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Do voting rights affect institutional investment decisions? Evidence from dual-class firms.

We examine whether, and to what extent, shareholder voting rights affect institutional investment decisions. We find that institutional ownership in dual-class firms is significantly lower than it is in single-class firms after controlling for other determinants of institutional investment. Although institutions of all types hold fewer shares of dual-class firms, this avoidance is more pronounced for long-term investors with strong fiduciary responsibilities than for short-term investors with weak fiduciary duties. Following the unification of dual-class shares into a single class, institutional investors increase their shareholdings in the unifying firm. Overall, our results suggest that voting rights are an important determinant of institutional investment decisions.

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"A dual-class stock structure, which carries unequal voting rights, is antithetical to the fair and fundamental principle of a 'one-share, one-vote' system and has no place in today's marketplace. Control of a corporation should come from owning a majority of shares, not owning special shares with special rights."

William D. Crist, former president of CalPERS Board of Administration (Business Wire, Sacramento, CA, April 21, 1999, "CalPERS Announces Investment Opinion on Nine of Corporate America's Poorest Financial and Economic Performers")

What determines the variation in institutional ownership across firms has received a great deal of attention in empirical research in financial economics. While previous work has shown that institutional investment is related to certain firm and stock characteristics, it is only recently that researchers have started to explore how corporate governance mechanisms affect institutional investment. Given that institutions are the largest class of investors in the US stock market and they have been successful in managing their clients' assets (Binay, 2005), understanding whether and how firms' corporate governance attributes affect their investment decisions is of great importance for the design of corporate governance.

In this paper, we empirically study institutional investor preferences for the stock of firms with an extreme form of governance, dual-class shares, where different share classes carry differential voting rights. (1) Since insiders of dual-class firms hold the majority of the shares with superior voting power, they are able to control the firm without holding large equity stakes and are largely isolated from external control pressures such as takeover threats. In general, outside investors can only purchase the shares with inferior voting rights. Gompers, Ishii, and Metrick (2007) indicate that in more than 70% of the dual-class firms, the shares with superior voting power are not traded. In cases where super voting shares are traded, these shares are largely held by insiders. As a result, in dual-class firms, outside investors have limited control rights even when their fractional ownership may give them substantial cash-flow rights. (2) In stark contrast, in single-class firms, each share carries one vote. Thus, outside investors' cash flow and control rights are identical.

Although important institutional investors often publicly voice their concerns about dual-class structures (see the quote above), a priori it is unclear whether institutional investment decisions should be affected by the lack of voting rights associated with outside equity positions in dual-class firms. On the one hand, dual-class share structures may not significantly affect institutional investment if institutions simply chase past returns, especially given that prior studies show no significant difference in performance between dual- and single-class firms. Moreover, the US possesses the best practice in security laws and corporate governance mechanisms. It is unlikely that outside shareholders of dual-class firms can be expropriated. On the other hand, institutions may hold less of the shares of dual-class firms due to the constraints arising from their prudence standards, the career concerns of portfolio managers, or their reduced ability to intervene. Hence, our analysis of institutional investment in dual-class firms sheds light on whether voting rights, which are arguably the most important type of shareholder rights (La Porta, Lopez-de-Silanes, Shleifer, and Vishny, 1997; Gompers et al., 2007), influence institutional investment decisions.

For both single-class and dual-class firms, we define institutional ownership as institutional investors' dollar investment in the firm's equity as a percentage of the firm's total market value of equity. We first explore whether institutional investment differs across single-class and dual-class firms. These cross-sectional tests demonstrate that aggregate institutional ownership in dual-class firms is about 3.6 percentage points lower than it is in single-class firms, after controlling for a host of factors that affect institutional investment in stocks. Although institutions of all types have lower ownership in dual-class firms than in single-class firms, the magnitude of the effect of a firm's dual-class status on institutional holdings varies across types. The effect is stronger for the group of long-term investors with important fiduciary responsibilities that are usually the most active in corporate governance, and it is weaker for short-term investors with low fiduciary responsibilities that are less likely to engage in shareholder activism.

We then use the time-series variation in firms' dual-class status to examine how the unification of dual-class structures into a single-class affects institutional investment. Our analysis demonstrates that relative to a control group of dual-class firms that do not unify their share classes, dual-class firms that unify their share classes experience a significant subsequent increase in institutional ownership. Moreover, after the unification takes place, all types of institutional investors increase their equity holdings in the unifying firms.

We also conduct a battery of additional tests to explore the robustness of our results. The analysis indicates that our findings are not driven by reverse causality. They are also robust to the use of various alternative measures of institutional ownership and alternative econometric specifications. Lastly, our main results are unaffected when we further control for insider ownership, board characteristics, and institutional herding.

Our paper contributes to the literature in the following ways. First, previous work finds that US institutional investment is related to stock and firm characteristics, as well as the quality of governance practices (Del Guercio, 1996; Gompers and Metrick, 2001; Bushee, Carter, and Gerakos, 2007). We add to this literature by demonstrating that US institutions invest substantially less in domestic firms with dual-class structures, where outside shareholders have little or no voting rights, even when the country-level investor protection and security laws are well developed.

Second, there is mixed evidence regarding whether dual-class arrangements hurt or increase firm value (Mikkelson and Partch, 1994; Dimitrov and Jain, 2006). We add to this literature by revealing that dual-class arrangements are associated with a diminished presence of institutional investors. Since institutions are the largest participants in the stock market, our findings suggest that the lack of shareholder voting rights may compromise dual-class firms' access to equity financing.

The paper is organized as follows. In Section I, we review the related literature, discuss the conceptual framework, and outline our empirical tests. Section II describes our sample formation and defines variables. Section III examines whether dual-class structures affect institutional ownership. In Section IV, we conduct several additional tests to explore the robustness of our findings and we present our conclusions in Section V.

I. Literature Review, Conceptual Framework, and Outline of Tests

A. Related Literature

1. Determinants of Institutional Ownership

Our paper fits within the growing literature that relates US institutional investment in domestic companies to stock and firm characteristics (Badrinath, Kale, and Ryan, 1996; Del Guercio, 1996; Falkenstein, 1996; Coval and Moskowitz, 1999,2001; Gompers and Metrick, 2001; Bennett, Sias, and Starks, 2003; Grinstein and Michaely, 2005). It is most closely related to a couple of recent studies that examine the relation between institutional investment and corporate governance. Bushee and Noe (2000) find that firms with higher disclosure ratings have greater institutional ownership. Bushee et al. (2007) demonstrate that institutional investors invest more in US firms with good board characteristics but are largely indifferent to the Gompers, Ishii, and Metrick (2003) Governance Index.

Our study is also related to the literature regarding how corporate governance affects the "home bias" in US investors' portfolio decisions. For example, Dahlquist, Pinkowitz, Stulz, and Williamson (2003), Leuz, Lins, and Warnock (2008), and Kho, Stulz, and Warnock (2006) reveal that US investors' portfolios under-weigh the stocks of foreign companies that are poorly governed. In addition, Aggarwal, Klapper, and Wysocki (2005) find that US mutual funds invest more in emerging markets with stronger country-level shareholder rights, legal frameworks, and accounting standards, as well as in those firms with better accounting disclosure.

More broadly, our work is also related to international and cross-country studies of the relation between institutional investment and governance attributes. In their study of Swedish firms, Giannetti and Simonov (2006) conclude that the majority of investors, including institutional investors, are reluctant to invest in companies with weak corporate governance. In a cross-country study, Giannetti and Koskinen (2008) demonstrate that institutional investors based in countries with poorer investor protection invest more abroad and put greater portfolio weights on countries with better investor protection.

2. Dual-Class Firms

Our study is also closely related to the literature concerning dual-class firms. Much of the literature examines whether dual-class arrangements hurt shareholder value, and the evidence is far from conclusive. Some authors argue that dual-class structures protect private control benefits (Smart and Zutter, 2003), and that the separation of ownership and control leads to value losses (Mikkelson and Partch, 1994; Claessens, Djankov, Fan, and Lang, 2002). Others, including Fama and Jensen (1983) and DeAngelo and DeAngelo (1985), contend that dual-class firms are not necessarily poorly managed due to the strong family involvement in these firms. Partch (1987), Cornett and Vetsuypens (1989), Lehn, Netter, and Poulsen (1990), Denis and Denis (1994), and Dimitrov and Jain (2006) provide evidence that dual-class structures do not destroy shareholder wealth. King and Segal (2008) show that Canadian firms with dual-class shares can increase their valuation by bonding themselves to the US securities regime through cross-listing. Bacon, Cornett, and Davidson (1997) demonstrate that the stock price reaction to dual-class recapitalizations is related to the board's characteristics. More recently, Gompers et al. (2007) explore the determinants of dual-class status and the performance consequences of differential voting and cash flow rights. They conclude that firm value is increasing in insiders' cash flow rights and decreasing in insiders' voting rights.

B. The Conceptual Framework

A priori, it is unclear whether the lack of voting rights to shareholders in dual-class firms should affect institutional investment decisions. There are several scenarios that suggest dual-class share structures may not significantly affect institutional investment. First, previous research has indicated that institutional investment decisions are primarily driven by past stock returns (Badrinath and Wahal, 2002; Nofsinger and Sias, 1999), and the empirical evidence regarding performance differences between dual and single-class firms is largely inconclusive (Partch, 1987; Cornett and Vetsuypens, 1989; Lehn et al., 1990; Denis and Denis, 1994; Bohmer, Sanger, and Varshney, 1996; Dimitrov and Jain, 2006).

Second, the US possesses very stringent security laws designed to protect shareholder rights and sets extremely high standards for corporate governance practices (La Porta et al., 1997; Giannetti and Koskinen, 2008). There is no evidence that minority or outside shareholders are expropriated in the US.

Third, although some important institutional investors publicly voice their concerns about dual-class structures, the extent to which such voting arrangements may affect their investment decisions is not obvious. For example, institutions cannot avoid stocks that are part of major market indices, and some stocks may be important for portfolio diversification. Thus, the aforementioned discussion suggests that institutional ownership in dual-class firms might not significantly differ from that in single-class firms.

On the other hand, there are arguments that support institutional managers' reticence to invest in the stock of dual-class companies, thus exhibiting a stronger preference for the stock of single-class firms than that of individual investors. First, unlike individual investors, institutional managers are subject to "prudence" standards that constrain their investment decisions (Del Guercio, 1996). To the extent that courts enforcing "prudent man" laws perceive investment in dual-class firms as exhibiting a low level of prudence, institutions may avoid these stocks to minimize their exposure to legal liabilities.

Second, given the negative view of dual-class structures shared by many important institutional investors, portfolio managers may avoid investing in the stock of dual-class firms to protect their careers. This is because in periods of lackluster performance, it is difficult for portfolio managers to demonstrate the soundness of their judgment to invest in dual-class firms. (3) Thus, a manager of a poorly performing portfolio with heavier emphasis on the stock of dual-class companies may be more likely to be dismissed than a manager whose portfolio places little weight on dual-class firms (Badrinath, Gay, and Kale, 1989). In contrast, individual investors' human capital is independent of the performance of their investment portfolios.

Third, institutions may prefer the stock of single-class firms over that of dual-class firms as they can use the voting power conveyed by their larger stakes to influence corporate decisions. This may be especially true when institutional selling of poorly performing stocks could be costly due to the potentially large impact on stock prices (see Smith, 1996; Carleton, Nelson, and Weisbach, 1998; Karpoff, 1998). In contrast, because of the high costs of direct action associated with their smaller stakes, individual investors, in general, are not actively involved in corporate governance and, as such, may place little value on voting rights. All of these scenarios suggest that institutions might choose to hold less of the stock of dual-class firms, which in turn implies that individual investors would choose to hold more of the shares of those firms. (4)

Given the conflicting views discussed above, whether institutional investors care about voting rights, thus avoiding the stock of companies with dual-class share structures, and to what extent remain open empirical questions.

C. Outline of Our Tests

Our first set of tests exploits the cross-sectional variation in firms' dual-class status. Using a sample of dual-class firms and the universe of single-class firms as the control group, we explore whether aggregate ownership by institutions differs across single-class and dual-class firms. We also examine whether the relation between firms' dual-class status and institutional ownership differs by type of institution.

Our second set of tests takes advantage of the time-series variation in firms' dual-class status to examine how the unification of dual-classes into a single class affects institutional ownership. We examine how institutional ownership changes after dual-class firms unify their multiple share classes into a single class, relative to a control group of dual-class firms that maintain their multiple-share-class structure.

II. Our Sample and Variable Definition

To form our sample, we start with the merged Center for Research in Security Prices (CRSP)-Compustat universe for the period 1995-2002. We focus on this period because we can accurately identify the dual-class firms each year using the data collected and generously made available to us by Gompers et al. (2007). These data, which we download from Andrew Metrick's website (http://www.som.yale.edu/faculty/am859/), is the most comprehensive US dual-class data set available. To our initial sample, we then merge institutional investors' holdings data from the Thomson Financial's CDA/Spectrum Database, stock market data from CRSP, accounting data from Compustat, and analyst coverage data from Institutional Brokers' Estimate System (IBES). (5) Our final sample consists of 614 dual-class firms (2,694 firm-year observations) and 8,360 single-class firms (37,503 firm-year observations) for the period 1995-2002. (6)

Our key variable of interest is institutional ownership in dual-class and single-class firms. In single-class firms, there is no difference between cash flow and voting rights. Hence, institutional ownership can be measured as the fraction of shares outstanding held by institutions or equivalently as the ratio of the dollar value of institutional investment to firm equity value. In dual-class firms, however, each share class carries differential cash flow and voting rights. Thus, institutional ownership of cash flow rights and of voting rights are different. We are interested in examining whether the level of institutional investment differs across single-class and dual-class firms. Therefore, the proper measure of institutional investment for dual-class firms is the dollar value of institutional investment in the firm as a percentage of firm equity value.

For both dual-class and single-class firms, we define institutional ownership (I0) as the fraction of a firm's equity value held by institutions measured in percentages. More precisely, institutional ownership in firm i with n different share classes outstanding (indexed by j) is constructed as follows:

[IO.sub.i] = [n.summation over (j=1)][P.sub.ji][s.sub.ji] / [n.summation over (j=1)][P.sub.ji][so.sub.ji], (1)

where [s.sub.ji] is the number of class j shares held by institutional investors in firm i, [so.sub.ji] is the total number of class j shares outstanding in firm i, and [P.sub.ji] is the class j share price of firm i. Note that institutional ownership in single-class firms is simply the fraction of shares outstanding held by institutions (n = 1). For dual-class firms with all classes of shares traded, we obtain the share price [P.sub.ji] from CRSP. For dual-class firms with nontraded superior-voting classes, we follow Gompers et al. (2007) and assume that the nontraded superior-voting shares have the same price as the traded inferior voting shares (i.e., a zero voting premium).

Panel A of Table I reports the mean and median institutional shareholdings (10) across dual-class firms and the universe of single-class firms for both the aggregate of institutional investors and by type of institution. The five types of institutions based on CDA/Spectrum's classification are: 1) bank trust departments, 2) insurance companies, 3) investment companies, 4) independent investment advisors, and 5) others. The institutions in this last group are a mix of ESOPs, university endowments, foundations, and private and public pension funds.

The table indicates that total institutional ownership is slightly higher in dual-class firms than in single-class firms, both at the mean and median, and the differences are statistically significant. The same pattern holds for the median holdings of all five types of institutions. The mean institutional ownership in dual-class firms is higher for bank trust departments, insurance companies, investment companies, and independent investment advisors, but lower for the group of other institutional investors. However, univariate statistics may mask the true relation between firms' dual-class status and institutional investment, especially if there are large differences in industry representation, as well as in firm and stock characteristics, across the two groups that are not controlled for.

In Panel B, we further compare institutional ownership in our sample of dual-class firms to that of a sample of single-class firms matched by industry and firm size. By matching each dual-class firm to a single-class firm of similar size in the same Fama-French industry, we remove some of the important differences in single and dual-class firms that affect institutional investment (Fama and French, 1997). This approach unveils that the mean (median) aggregate institutional ownership in dual-class firms is about 3.24 (3.12) percentage points lower than that in similar single-class firms of the same industry. This difference is statistically significant at the 0.01 level. Further, institutions of every type hold less of the shares in dual-class firms than they do in similar single-class firms, except for insurance companies where the median holdings are slightly higher for dual-class firms. Thus, the evidence in Panel B suggests that after controlling for industry membership and firm size, institutional ownership in dual-class firms is lower than in single-class firms.

In our multivariate analysis, we use institutional ownership (IO) as the dependent variable. Our key test variable is a firm's dual-class status, the Dual dummy, which equals one if the firm has multiple share classes, and zero otherwise. Our control variables comprise firm and stock characteristics that previous research has shown to determine institutional investment. Market capitalization, Mktcap, is defined as the dollar value of all share classes at the end of the year, in millions of 2002 dollars. The annual return on the firm's stock, Return, is defined as the value-weighted average of the returns across traded classes over the year. The dividend yield, Divyield, is defined as the ratio of total dividend payout to stock price. The volatility of stock returns, Retvol, is defined as the value-weighted average of the stock return volatility across traded classes using monthly stock returns over the prior year. The share turnover ratio, Turnover, is defined as the value-weighted average of the ratio of the trading volume to the number of shares outstanding at the end of the previous year across all traded classes. The market-to-book ratio, M/B, is defined as the market value of assets divided by the book value of assets. Financial leverage, Leverage, is computed as the ratio of total debt to the market value of assets. Firm age, Firmage, is defined as the number of years since the firm first appears in CRSP. Share price, Price, is defined as the value-weighted average of the stock price across traded classes at the end of the year, in 2002 dollars. S&P 500 membership, S&P 500, is a dummy equal to one if the firm is in the S&P 500 index, and zero otherwise. Analyst coverage, #Analysts, is defined as the number of analysts coveting the firm according to IBES. Firm age, S&P 500 membership, and analyst coverage are proxies for aggregate information and/or the cost of information collection. (7) Share price and turnover capture transaction costs. Table II presents summary statistics for these variables.

It is clear that firm and stock characteristics differ substantially between dual-class and single-class firms, except for stock returns. The median market capitalization of dual-class firms is significantly larger than that of single-class firms, while the average market capitalization of single-class firms is significantly larger than that of dual-class firms. The average dividend yield of dual-class firms is larger than that of single-class firms. Dual-class firms appear to have significantly lower return volatility, lower turnover, lower market-to-book ratios, higher leverage, higher share price, lower likelihoods of being part of the S&P 500 index, and lower analyst coverage. Thus, results in Table II suggest that, in addition to controlling for industry membership and firm size, it is important that we control for an extensive list of potential determinants of institutional investment in our multivariate analysis.

III. Institutional Investment in Dual-Class Firms

Before proceeding with our multivariate analysis, we examine the correlation between our right-hand-side variables. Table III indicates that the extent of correlation among most pairs of variables raises little concern for multicollinearity in our regression analysis. There are, however, a few moderate correlations in the order of 0.30-0.40 (e.g., between dividend yield and firm age) with a maximum of 0.52 between analyst coverage and the S&P 500 membership dummy. Nevertheless, we note that the results from our multivariate analysis are robust to different specifications that exclude some of the right-hand-side variables in each of the moderately correlated pairs.

Throughout our regression analysis, we control for industry effects using 48 Fama-French industry dummies. This approach ensures that what identifies our estimated coefficients is the cross-sectional variation in dual-class status within firms in the same industry, removing any time-invariant industry-specific characteristics that could drive the results. In addition, we include year dummies to control for changing market conditions or trends that may affect institutional investment over time. To assess the statistical significance of our results, throughout the analysis we use Rogers' (1993) robust standard errors that adjust for the clustering of observations at the firm level by assuming that observations are independent across firms, but not within firms.

Although we consider the pooled ordinary least squares (OLS) regressions with clustered standard errors to be the most appropriate in our context, for robustness we also report the results of Fama-MacBeth regressions (Fama and MacBeth, 1973) with Newey-West standard errors (Newey and West, 1987). We caution, however, that this method may be less reliable given the few years of observations in our sample period.

A. Aggregate Institutional Investment in Dual- versus Single-Class Firms

To explore whether voting rights are an important consideration in institutional investment decisions, we regress institutional ownership (IO) on the dual-class status dummy (Dual) and the set of firm and stock characteristics defined above. The coefficient on Dual captures the difference in institutional holdings of dual- versus single-class firms, after controlling for other differences across these two types of firms. If voting rights have no effect on institutional investment decisions, we expect the coefficient on Dual to be insignificantly different from zero. We expect the coefficient to be negative if the lack of voting rights discourages institutional investment. Table IV reports the results.

In Panel A, we estimate pooled OLS regressions with clustered standard errors. Across all four specifications, we find a negative and statistically significant coefficient on the Dual dummy. The magnitude of the effect is not only statistically significant, but also economically important. The estimates in Column 4, where all control variables are included, imply that aggregate institutional ownership in dual-class firms is 3.6 percentage points lower than it is in single-class firms. Given that the average fractional holding of institutional investors across all firms in the sample is about 33 percentage points over our sample period, this difference is economically significant. Institutional ownership is 11% lower in dual-class firms than in single-class firms. Thus, we find evidence that institutional investors tend to invest less in dual-class firms. In terms of the control variables, most of our findings conform to those in prior studies. Institutions invest more in larger, older firms, firms with lower prior year returns, lower dividend yields, lower return volatility, higher turnover, lower market-to-book ratios, higher leverage, higher stock prices, and greater analyst coverage.

Panel B reports the results from Fama-MacBeth regressions with Newey-West standard errors based on five lags. All regressions include the same control variables as in Panel A. We omit reporting coefficients on the control variables for brevity. We note that the results on the effect of dual-class status on institutional investment are similar to those under the pooled OLS specifications.

Roughly, one-third of the dual-class firms in our sample have multiple traded classes. When shares with inferior voting power (as well as those with superior voting power) are traded, institutional investors can, in principle, purchase some of the superior-voting shares. However, this is uncommon since the firm's insiders usually hold the superior-voting shares even when they are traded (Gompers et al., 2007). Nonetheless, it is possible that institutional investors may seek to purchase such shares to obtain a voice in corporate matters. As a result, dual-class firms with both classes traded may be relatively more attractive to institutional investors than those where only the ordinary shares with little or no voting power are traded. To explore this possibility, we repeat our analysis of the effect of dual-class status on institutional ownership by excluding dual-class firms where both classes are traded. Panel C reports the results.

We find that the negative effect of dual-class status on institutional ownership is stronger in this subsample. Institutional ownership in dual-class firms with nontraded superior-voting shares is about 4.7 percentage points lower than in single-class firms (Column 4). Thus, the evidence is consistent with the view that institutional investors choose to hold even fewer shares of dual-class firms when the superior-voting shares are not traded as they find it impossible to exert any influence on firms' decisions.

Since ownership by institutions and ownership by individuals must add up to 100%, an equally valid interpretation of our results is that the aggregate ownership by individual investors is higher in dual-class firms than in single-class firms. Thus, any potential explanation of our results must not only explain why the stock of dual-class firms is less attractive to institutional investors but also why these securities are relatively less attractive to institutional investors than to individual investors. As discussed in Section I.B., institutional managers may place a lower value on the stock of dual-class firms than do individual investors due to their fiduciary duties, career concerns, and their ability to intervene. Thus, individual investors are willing to hold a larger fraction of a dual-class firm's equity while institutional investors are willing to hold a smaller portion.

To summarize, the collective evidence from Table IV suggests that the aggregate institutional holdings in dual-class firms are significantly lower than those in single-class firms, after accounting for other factors that affect institutional investment. These results provide strong evidence that shareholder voting rights do affect institutional investment decisions, despite the fact that the US has the most stringent corporate governance requirements and offers the greatest protection to shareholders. The aggregation of institutional holdings, however, may mask important heterogeneity across different types of institutional investors (Brickley, Lease, and Smith, 1988; Del Guercio, 1996; Woidtke, 2002; Chen, Harford, and Li, 2007). We next explore whether voting rights affect the investment decisions of different types of institutions in different ways.

B. Institutional Investment by Type of Institution

Differences in institutional investment across types of institutions may arise as a result of variations in their fiduciary responsibilities, investment horizon, objectives, or styles. Also, different types of investors may have diverse assessments of shareholder voting rights depending on whether they are more or less likely to engage in shareholder activism.

Investment companies and independent investment advisors are usually short-term investors that rebalance their portfolios often, have low levels of fiduciary responsibility, and do not engage in shareholder activism. Thus, these investors are likely to be the least sensitive to voting rights. Conversely, long-term investors with strong fiduciary responsibilities, who are more likely to engage in shareholder activism, such as pension plans and university foundation endowments, are likely to be highly sensitive to shareholder voting rights. It is less clear whether voting rights should matter in the investment decisions of bank trust departments and insurance companies. Both types of institutions are long-term investors, and bank trust departments further have strong fiduciary duties, suggesting that shareholder voting rights should matter in their investment decisions. However, both types of institutions also have important potential business relations with the firms they invest in, and, as such, may not use their voting rights against management. This suggests that voting rights might be of little value to them.

To investigate whether a firm's dual-class status affects institutional investment differently across different institution types, in Table V we regress institutional ownership by each type of institution on Dual and the same control variables as in Table IV.

Panel A reports the pooled OLS regressions with clustered standard errors, while Panel B displays the results using the Fama-MacBeth regressions with Newey-West standard errors based on five lags. The table indicates that all types of institutional investors appear to invest less in dual-class firms. The results are similar across different estimation methods, except for bank trust departments where the coefficient on the Dual dummy is negative under both specifications, but only statistically significant using the Fama-MacBeth procedure.

As before, for each institution type, the coefficient on Dual captures the difference in institutional holdings of dual versus single-class firms, after controlling for firm and stock characteristics. To assess the economic magnitude of these differences and the relative importance across investor types, we normalize the coefficient on Dual reported in Panel B by the average ownership in single-class firms by each institution type (from Panel A, Table I). This calculation illustrates that bank trust departments have ownership stakes in dual-class firms that are 7.2% lower than in single-class firms. The measure is about 16.9% for insurance companies, 10.6% for investment companies, 10.3% for independent investment advisors, and 17.3% for the other investors type.

The effects of dual-class status for insurance companies and the other investors are not statistically different from each other. The difference in the effect of dual-class status for investment companies and independent advisors is statistically significant, but not economically significant. The effect for both insurance companies and the group of other institutions is statistically and economically higher than for investment companies and independent advisors. The effect for investment companies and independent advisors is statistically and economically higher than for bank trust departments. Thus, the group of other investors, which includes the most important shareholder activists, together with insurance companies, are the types of investors with the lowest (relative) investment in dual-class firms, followed next by independent investment advisors and investment companies, followed by bank trust departments.

As discussed earlier, we cannot rule out the possibility that institutional investors may seek to intervene in corporate matters by purchasing shares with superior voting power when such shares are traded. In Panel C, we repeat our pooled OLS regression analysis of the effect of dual-class status on institutional ownership by excluding dual-class firms where both classes are traded. As before, standard errors are clustered at the firm level. We find that for each type of institution, the negative effect of dual-class status on institutional ownership is statistically significant, and sometimes even stronger, in this subsample than that reported in Panel A.

To summarize, we find that institutional investors tend to invest less in the stock of dual-class firms, and that the effect of firms' dual-class status on their investment decisions is largely independent of institutional manager types. Our results suggest that voting rights are an important consideration for institutional investors when making their portfolio decisions and that those with more stringent fiduciary responsibilities and longer investment horizons, as well as those more commonly associated with shareholder activism, are more sensitive to the lack of voting rights in dual-class firms.

C. Changes in Institutional Ownership Following Unification

To shed further light on whether a firm's dual-class status affects institutional investment decisions, in this subsection we examine how the unification of a dual-class structure into a single class affects institutional investment. The sample for our unification analysis includes both dual-class firms that remain so for the entire sample period and those that abandon their dual-class structures up to one year after the unification. The sample contains 79 unification events and 2,160 firm-year observations. We examine changes in institutional ownership following the elimination of dual-class structures, using the remaining (nonunifying) dual-class firms as the control group. In this way, our analysis captures the changes in institutional ownership due to unification over and above changes motivated by reasons that are common across dual-class firms.

In Panel A of Table VI, we regress the change in the level of institutional ownership (ChgIO) from year t to t + 1 (i.e., ChgIO = [IO.sub.t+1] - [IO.sub.t]) on the Unify dummy, which is equal to one if the firm abandons its dual-class structure in year t, and zero otherwise. The standard errors of the coefficients are clustered at the firm level. The control variables are measured in levels as of year t and are identical to those previously defined. As an additional control, we include ChgShouts, the percentage change in the total number of shares outstanding from year t to year t + 1. This variable captures any new equity issues or repurchases following the unification that may affect the change in institutional ownership. It also controls for any effect that the exchange of shares as a result of the unification may have on institutional ownership.

Following the unification of a dual-class structure into a single class, there is a large increase in the institutional ownership of the unifying firms over and above the change in institutional ownership experienced by the control group of nonunifying dual-class firms. The unification is associated with a 10.8 percentage point increase in total institutional ownership (see Column 1 of Panel A). When compared to the preunification fractional ownership by institutions (almost 35 percentage points), this implies a 30.9% increase.

The analysis by type of institution in Columns 2-6 of Panel A demonstrate that, following the unification of a dual-class structure, all types of institutional investors significantly increase their holdings in the unifying firms relative to the control group of nonunifying dual-class firms. In addition, the increase in institutional holdings of the unifying firms is economically significant. When compared to the preunification fractional ownership, the estimates in Panel A imply a 29.7% increase for bank trust departments, 43.5% for insurance companies, 43.2% for investment companies, 23.1% for independent investment advisors, and 38.6% for other institutional investors. These findings suggest that most institutional investors are seriously concerned about the poor corporate governance in dual-class firms and that they significantly increase their investment after the dual-class structure is removed.

In Panel B, we repeat the analysis in Panel A, but we measure the control variables in changes between year t and year t-1 instead of in levels at year t. These changes are indicated by the symbol A preceding the corresponding variable. This specification addresses the concern that changes in the control variables, such as changes in liquidity or other factors, could be correlated with the unification event and at the same time cause changes in institutional ownership. If this is the case, then the estimated coefficient on Unify reported in Panel A could be biased. However, our results indicate otherwise as replacing the controls in levels by those in changes has little effect on the coefficient on Unify.

It is possible that our dependent variable, ChgIO, and, thus our inference reported in Panels A and B, may be contaminated by the mechanics of the unification process. As part of the process, the superior-voting shares are exchanged for shares of the surviving class. As a result, both the total number of shares outstanding (the denominator of IO) and the number of shares held by institutional investors (the numerator of IO) could be affected.

The numerator of IO, and, consequently ChgIO, could be measured with error if institutional investors hold superior-voting shares that are exchanged for common shares. This problem would be minimized in firms where the superior-voting shares are not traded. Thus, institutional holdings of these shares are likely to be negligible. Therefore, we repeat our analysis on the subsample of dual-class firms where only the ordinary shares are traded. In Panel C, we include the control variables in levels and in Panel D, we include the control variables in changes. The results remain statistically significant and qualitatively similar to those reported in Panels A and B.

Our inference might also be affected due to the change in the number of shares outstanding that affects the denominator of our institutional ownership measure (IO). To explore this issue, we construct a new variable, %ChgIIShrs, defined as the percentage change in the number of shares held by institutional investors, and use it in place of ChgIO as the dependent variable. (8) Since %ChgIIShrs does not require the number of shares outstanding for its calculation, it will not be affected by any change in the total number of shares outstanding due to unification. Table VII reports the results. In Panel A, the control variables are in levels, and in Panel B, the control variables are in changes. The results from both panels indicate that aggregate institutional ownership increases following the unification event. With the exception of only insurance companies, there is statistically significant evidence that institutions of all types increase their shareholdings post unification.

In Panels C and D of Table VII, we repeat the analysis using%ChgIIShrs as the dependent variable but limit the sample to dual-class firms where only the ordinary shares are traded. In this case, the mechanics of the unification cannot affect our inference at all. %ChgIIShrs is not affected by the change in the total number of shares outstanding, and institutional investors do not participate in the share exchange because they hold either none or very little nontraded superior-voting shares prior to the unification. Again, our main inferences based on the results in Panels A and B remain unchanged regardless of whether the control variables are measured in levels or in changes. (9)

To summarize, we find that unifications are followed by large increases in institutional ownership, and that the mechanics of the unification process are unlikely to drive our results. (10) Thus, the evidence from unifications is consistent with our previous evidence that institutional investors do care about voting rights in making their investment decisions.

IV. Additional Investigation

A. Are Our Results Due to Reverse Causality?

In our empirical analysis, we model institutional investors' investment decisions as a function of firms' dual-class status, which we treat as an exogenous stock characteristic. However, Bushee et al. (2007) demonstrate that the level and changes in ownership by governance sensitive institutional investors are associated with future changes in governance. Thus, a natural concern that arises is whether it is possible that causality could go in the other direction. A firm's decision to keep or abandon its dual-class structure may be a function of institutional investment decisions. In particular, the negative relation between dual-class status and institutional ownership that we document in Tables IV and V could arise if institutions with larger stockholdings are able to force firms to abandon their dual-class structures. (11)

To address the concern of reverse causality, we examine whether the level and changes in institutional ownership are associated with future unifications of dual-class structures into a single class using the approach in Bushee et al. (2007) in Table VIII. For this purpose, we focus on a sample that contains two types of dual-class firms: 1) those that unify their share classes until the year of unification and 2) those that maintain their multiple share classes during the entire sample period. We use three different specifications of probit models to examine the effect of institutional ownership on the probability of future unification. In Panel A, we estimate the probability as a function of lagged institutional ownership and control variables. In Panel B, we estimate the probability as a function of lagged institutional ownership and contemporaneous changes in the control variables. In Panel C, we estimate the probability as a function of lagged changes in institutional ownership and contemporaneous changes in the control variables. As noted by Bushee et al., including the contemporaneous changes in the control variables in the last two panels ensures that our results are not driven by an omitted relation between the level of (or changes in) institutional ownership and future changes in the control variables.

Our main test variables are total institutional ownership, IO, defined as institutional investors' dollar investment in the firm's equity as a percentage of the firm's total market value of equity; the ownership in the firm by bank's trust departments, BTIO; the ownership by insurance companies, INSCIO; the ownership by investment companies, INVCIO; the ownership by independent advisors, IAIO; and the ownership by the other types of institutions, OIIO. Among the control variables, #Mergers is the number of mergers in the firm's Fama-French industry, Capex is capital expenditures divided by assets, and ROA is the firm's return on assets. Distress is a dummy equal to one if the firm's cash flow is less than the firm's interest payments, and zero otherwise, Retearn is the ratio of the firm's retained earnings to total assets, and #Shares is the total number of shares outstanding across all share classes. All other variables are as previously defined. In Panels B and C, the symbol A preceding a variable denotes its change between year t and year t-1. All regressions include year and 48 Fama-French industry dummies (not reported) and adjust for the clustering of observations at the firm level.

The table indicates that the prior level of institutional ownership and the prior changes in institutional ownership have no statistically significant effect on the probability that the firm will unify its share classes in the following year. This result holds for both the aggregate institutional ownership variables as well as for the institutional ownership variables based on each type of institution. These findings, together with the evidence in Section III.C that institutions increase their ownership in the firm following unifications, suggest that reverse causality is not the reason behind the negative relation between dual-class status and institutional ownership we document in Sections III.A and III.B. Thus, we conclude that our evidence is consistent with institutions choosing to invest less in dual-class firms than in single-class firms.

B. Insider Ownership and Board Characteristics

Dual-class firms are often family firms, where insiders typically hold a large fraction of the outstanding shares. Therefore, it is possible that institutional ownership is systematically lower in dual-class firms simply because higher insider ownership reduces the fraction of shares available for outside investors to purchase. In addition, prior research indicates that institutional investors exhibit preferences for good board characteristics (Bushee et al., 2007). If these board characteristics are correlated with dual-class status, then their omission from the regression analysis could bias our results. For instance, if dual-class firms also have bad board characteristics, the negative effect of Dual on institutional ownership could be driven by the omitted board characteristics.

To explore these possibilities, we match our data with Institutional Shareholder Services/ Investor Responsibility Research Center (ISS/IRRC) data on the equity ownership of corporate insiders and board characteristics. Although this merge substantially reduces our sample size, we repeat our analysis in Table IV after controlling for insider ownership, InsiderOwn, defined as the percentage of the firm's equity held by insiders (executives and directors). BoardSize is the number of directors serving on the board, PctIndep is the percentage of board members that are independent directors, and CEOCOB is a dummy equal to one if the CEO is also the chairman of the board, and zero otherwise. The results are reported in Table IX.

Column 1 indicates that the magnitude of the negative effect of Dual on IO is much larger in this subsample (about 9.6 percentage points), even before controlling for insider ownership and board characteristics. Columns 2-5 add each of InsiderOwn, BoardSize, PctIndep, and CEOCOB, respectively, and Column 6 includes all the variables simultaneously. We find that even after controlling for insider ownership and board characteristics, the negative effect of dual-class status on institutional ownership remains statistically significant. Consistent with our expectations, higher insider ownership is associated with lower institutional investment, and good board characteristics (smaller and more independent boards) are associated with higher institutional ownership. We conclude that the lower institutional ownership in dual-class firms is not spuriously driven by a correlation between dual-class status and insider ownership or board characteristics.

C. Institutional Herding

Prior work argues that institutional investors often herd in their investment decisions and, thus, in the level of ownership they establish in any given firm (Nofsinger and Sias, 1999). If institutional herding is correlated with a firm's dual-class status, then this could bias our inference due to the classical omitted variables problem.

To explore whether our results are robust to this concern, we further control for institutional herding using the Lakonishok, Shleifer, and Vishny (1992) measure, commonly used in the literature, in Table X. The herding variable, which we denote as Herd, is constructed to measure the imbalance of institutional trading between purchases and sales. For each institution and each stock in year t, we first determine the change in the number of shares held from year t-1 to year t, adjusted for stock splits. We code each institution as being a net buyer if the change is positive and a net seller if it is negative. Herding by institutions for each stock in year t, Herdt, is defined as the number of net buyers divided by the sum of the number of net buyers and the number of net sellers. Panel A reports the coefficients from OLS regressions with clustered standard errors for both aggregate institutional ownership and for each type of institution. Panel B repeats the analysis using the Fama-MacBeth procedure with Newey-West standard errors.

The coefficient on Herd is positive and statistically significant both in our regressions for aggregate institutional ownership and in those for each type of institution. More importantly, in all columns, the coefficient on Dual remains negative and statistically significant, and slightly increases in magnitude as compared to those reported in Tables IV and V. Moreover, the coefficients on the control variables also remain similar to those reported before. Thus, we conclude that institutional herding does not drive our results.

D. Institutional Ownership under Alternative Assumptions

When the shares with superior-voting rights in dual-class firms are not traded, the calculation of our institutional ownership variable (IO) assumes that these shares have the same price as the traded ordinary shares (i.e., a zero voting premium). To the extent that institutions hold very few of the superior-voting shares, the assumption of a positive voting premium would have little effect on the numerator of our IO measure (the estimated value of institutional holdings in the firm). However, a higher voting premium will likely increase the denominator (the market value of firm equity) and, as a result, would reduce our measure of institutional ownership (IO) for dual-class firms with nontraded superior-voting shares.

Zingales (1995) indicates that the mean (median) premium for the superior-voting class relative to the inferior voting class is 10% (3%), so we explore how different assumptions on the voting premium affect our main analysis. In Table XI, we report the regression results using alternative assumptions for the voting premium.

In Panel A, we assume a 5% voting premium in the calculation of IO, and in Panel B, we assume a 10% voting premium. The tables demonstrate that, consistent with our conjecture, the higher voting premium is associated with a larger negative effect of Dual on IO. Recall that the estimates in Column 4 of Table IV imply that institutional ownership is 3.6 percentage points lower in dual-class firms than in single-class firms. Results in Table XI indicate that this difference is almost 4.0 percentage points when the voting premium is assumed to be 5% (Column 1 of Panel A), and 4.4 percentage points when the premium is assumed to be 10% (Column 1 of Panel B). We find similarly larger effects by type of institution.

E. Using an Alternative Sample Based on a Matching Procedure

The analysis reported in Tables IV and V compares institutional ownership in dual-class firms to that in the universe of single-class firms. As a robustness check, we repeat our analysis on the sample based on the matching procedure discussed in Section II, where we match each dual-class firm in our sample to a single-class firm based on 48 Fama-French industry and total assets. Table XII reports the multivariate regression analysis using IO as the dependent variable. Panel A reports the pooled OLS regressions with clustered standard errors, and Panel B reports the coefficients from the Fama-MacBeth regressions with Newey-West standard errors based on five lags.

The OLS results demonstrate that aggregate institutional ownership is about 3.3 percentage points lower in dual-class firms than in single-class firms (Column 1 of Panel A), and this difference is similar if we use the Fama-MacBeth procedure (Column 1 of Panel B). Both panels show a uniform negative effect of dual-class status on institutional ownership by type (Columns 2-6), except for bank trust departments. Overall, the results are statistically significant and of similar magnitude to those reported in Tables IV and V that use the universe of single-class firms as the control group.

V. Conclusion

We find that institutional ownership in dual-class firms is substantially lower than it is in comparable single-class firms, and this result holds for all types of institutions. This difference in investment is more pronounced for long-term investors with strong fiduciary responsibilities than for short-term investors with low levels of fiduciary responsibility. In addition, following the unification of dual-class structures, institutions substantially increase their investment in the new single-class firms. We conclude that the vast majority of institutional investors do avoid investing in the stock of dual-class firms. Moreover, this avoidance is economically significant.

Overall, we demonstrate that the lack of shareholder voting rights in dual-class firms is an important factor in institutions' portfolio decisions, suggesting that a firm's corporate governance attributes matter for institutional investment. In addition, although dual-class share structures may have some benefits for firms, such as allowing management to focus on long-term value without worrying about potential takeovers, our study suggests that they may also compromise firms' access to equity capital by discouraging investment by institutional investors.

We thank Andrew Metrick and Joy Ishii for sharing their dual-class data set. We thank Bill Christie (the editor), an anonymous referee, Xia Chen, Qiang Cheng, Martin Kacperczyk, Michael King, Karthik Krishnanm, and participants at the Northern Finance Association (NFA) Meeting in 2007 for comments and helpful suggestions. Priscille Aeschlimann, Huasheng Gao, Dermot Murphy, Dave Newton, and Bing Yu provided excellent research assistance. At the 2007 NFA Meetings, this paper was the winner of the Best Paper Award in valuation sponsored by the Canadian Institute of Chartered Business Valuators. We gratefully acknowledge the financial support from the Social Sciences and Humanities Research Council of Canada, the Bureau of Asset Management of the Sauder School of Business, and the Certified Management Accounting Society of British Columbia. All errors are ours.

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(1) Dual-class firms constitute about 6% of all firms and 8% of the total market capitalization in the United States (Gompers et al., 2007). There is also an increasing trend in dual-class IPOs, which account for 9% of all IPOs in 2005 as opposed to 4.5% in 1995 based on data from the SDC.

(2) Gompers et al. (2007) document that outside investors in dual-class firms hold about 60% of the cash flow rights but only 40% of the voting rights.

(3) Influential institutional investors, such as CalPERS and TIAA-CREE as well as governance rating and proxy voting services, all publicly denounce the dual-class structure.

(4) This would occur even when both individual and institutional investors may have some preference for the stock of single-class firms over that of dual-class firms. This is because the share price of a dual-class firm will not fully discount the lack of voting rights to outside shareholders, precisely since the equilibrium stock price reflects the demand of institutional and individual investors with heterogeneous valuations of the stock of dual-class firms. Thus, at the equilibrium stock price, institutional investors with a lower valuation will hold less of the shares of dual-class firms than individual investors with a higher valuation.

(5) Under rule 13(f) of the 1978 amendment to the Securities and Exchange Act of 1934, all institutions with greater than $100 million of equity securities under discretionary management are required to report their holdings quarterly. Common stock positions greater than 10,000 shares or $200,000 must be disclosed.

(6) While dual-class structures tend to concentrate in some industries, such as alcoholic beverages, printing and publishing, and telecommunications, they are not necessarily an industry-specific phenomenon. In fact, dual-class firms are present in 41 of the 48 Fama-French industries (Fama and French, 1997).

(7) For example, older stocks have a more established reputation and, thus, less estimation uncertainty of the riskiness of the stock. S&P 500 membership adds visibility to the stock. Greater analyst following serves as a proxy for the recent amount of useful information on the firm.

(8) The number of observations is smaller in Table VII than that in Table VI because the variable% ChgIIShrs is not defined when institutional ownership in year t is zero. The results are similar if the dependent variable is defined as the absolute change in the number of shares held by institutional investors (number of shares held by institutions in year t + 1 minus number of shares held by institutions in year t).

(9) We also used the change in the number of institutional shareholders as the dependent variable, which is unaffected by the unification process. We find that although the coefficient on Unify is always positive, it is only statistically significant in the subsample containing dual-class firms with only the ordinary classes traded when the dependent variable is the change in the total number of institutions, the number of bank trust departments, or the number of investment companies. In summary, the results are weaker, likely because the change in the number of institutions is a less precise measure of the change in the value of institutional investments in the firm, but they are largely consistent with those reported.

(10) Moyer, Rao, and Sisneros (1992) find that dual-class recapitalizations are followed by increases in institutional investment in their sample of 114 firms that recapitalized from 1979-1987. This would imply a positive relation between dual-class status and institutional ownership. Our findings differ from theirs largely because the tests correspond to very different sample periods. In particular, institutional investors' perception of the merit of dual-class shares may have been reversed in our more recent sample period relative to the 1980s. This is evidenced by the public denouncement of dual-class structures by some important institutional investors, such as CalPERS and TIAA-CREF, as well as governance rating and proxy-voting services (e.g., the Institutional Shareholder Services (ISS), the Council of Institutional Investors, the Governance Metrics International, and the Corporate Library).

(11) Reverse causality is less of a concern in our time-series tests in Section III.C, since we demonstrate that following the unification of dual-class structures into a single class, institutional investors increase their shareholdings in the unifying firm.

Xinlei Zhao, Kai Li is the W.M. Young Professor of Finance at the Sauder School of Business, University of British Columbia, BC, Canada. Hernan Ortiz-Molina is an Assistant Professor of Finance at the Sauder School of Business, University of British Columbia, BC, Canada. Xinlei Zhao is an Associate Professor of Finance at Kent State University, OH.
Table I. Institutional Ownership in Single-Class versus Dual-Class
Firms

The sample contains 8,360 single-class firms with a total of 37,503
firm-year observations and 614 dual-class firms with a total of 2,694
firm-year observations for the period 1995-2002. Institutional
ownership is defined as institutional investors' dollar investment in
the firm's equity as a percentage of the firm's total market value of
equity. We report mean values and median values in parentheses below.
The last column reports the p-value for t-tests of the difference in
means and Wilcoxon rank-sum tests for the difference in medians (in
parentheses). Panel A reports institutional ownership in single-class
and dual-class firms in our sample. Panel B reports institutional
ownership in dual-class firms and in a control sample of single-class
firms matched by 48 Fama-French industry and total assets.

 Single- Dual-
 Class Class Difference p-value

Panel A. Institutional Ownership-Full Sample

Institutional ownership 33.23 34.74 1.52 0.006
 (28.35) (33.55) (5.20) (0.000)
Bank trust departments 4.22 4.69 0.47 0.000
 (2.00) (2.93) (0.93) (0.000)
Insurance companies 2.25 2.26 0.01 0.829
 (0.65) (1.23) (0.58) (0.000)
Investment companies 8.09 8.40 0.31 0.114
 (3.81) (6.04) (2.23) (0.000)
Independent investment 15.90 16.78 0.87 0.001
 advisors (13.52) (15.27) (1.75) (0.000)
Other institutional 2.83 2.64 (0.19) 0.036
 investors (0.75) (1.30) (0.55) (0.000)

Panel B. Institutional Ownership-Matched Sample

Institutional ownership 37.98 34.74 (3.24) 0.000
 (36.67) (33.55) 0.00 0.002
Bank trust departments 4.82 4.69 (0.13) 0.426
 (2.89) (2.93) 0.00 0.268
Insurance companies 2.70 2.26 (0.44) 0.001
 (1.21) (1.23) (0.02) 0.203
Investment companies 9.04 8.40 (0.65) 0.012
 (6.09) (6.04) 0.00 0.232
Independent investment 18.37 16.78 (1.59) 0.000
 advisors (17.19) (15.27) 0.00 0.006
Other institutional 3.11 2.64 (0.48) 0.000
 investors (1.51) (1.30) 0.00 0.814

Table II. Sample Characteristics

The sample contains 8,360 single-class firms with a total of 37,503
firm-year observations and 614 dual-class firms with a total of
2,694 firm-year observations for the period 1995-2002. We report
mean values and median values in parentheses below. The last column
reports the p-value for t-tests of the difference in means and
Wilcoxon rank-sum tests for the difference in medians (in
parentheses). Mktcap is market capitalization, defined as the
dollar value of all share classes at the end of the year (in
millions of 2002 dollars); Return is the annual return on the
firm's stock, defined as the value-weighted average of the returns
across traded classes over the year; Divyield is the dividend
yield, defined as the ratio of total dividend payout to stock
price; Retvol is the volatility of stock returns, defined as the
value-weighted average of the stock return volatility across traded
classes using monthly stock returns over the prior year; Turnover
is the share turnover ratio, defined as the value-weighted average
of the ratio of the trading volume to the number of shares
outstanding at the end of the previous year across all traded
classes; M/B is the market-to-book ratio, defined as the market
value of assets divided by the book value of assets; Leverage is
financial leverage, defined as the ratio of total debt to the
market value of assets; Firmage is firm age, defined as the number
of years since the firm first appears in CRSP; Price is the share
price, defined as the value-weighted average of the stock price
across traded classes at the end of the year (in 2002 dollars); S&P
500 is an S&P 500 membership dummy, which equals one if the firm is
in the S&P 500 index, and zero otherwise; and #Analysts is analyst
coverage, defined as the number of IBES analysts covering the firm.

 Single- Dual-
 Class Class Difference p-value

Mktcap (market 2377.03 1903.72 (473.31) 0.090
 capitalization) (152.91) (331.78) (178.87) (0.000)
Return (annual return) 0.20 0.19 -0.01) 0.543
 (0.01) (0.04) (0.03) (0.002)
Divyield (dividend yield) 0.66 0.81 0.15 0.000
 (0.00) (0.00) (0.00) --
Retvol (return volatility) 0.64 0.52 -0.12 0.000
 (0.53) (0.43) (-0.10) (0.000)
Turnover (shares turnover) 14.76 9.17 (5.59) 0.000
 (9.10) (5.62) (-3.48) (0.000)
M/B (market-to-book) 2.03 1.52 -0.51 0.000
 (1.18) (1.03) (-0.15) (0.000)
Leverage (financial 0.22 0.29 0.07 0.000
 leverage) (0.14) (0.22) (0.08) (0.000)
Firmage (age since 12.48 13.52 1.05 0.000
 listing) (7.00) (9.00) (2.00) (0.000)
Price (share price) 20.69 180.87 160.18 0.000
 (10.92) (16.91) (5.99) (0.000)
S&P500 (S&P 500 0.08 0.07 (-0.01) 0.065
 membership) (0.00) (0.00) (0.00) --
#Analysts (analyst 1.53 1.44 -0.09 0.059
 coverage) (0.58) (0.67) (0.09) (0.001)

Table III. Correlation Matrix

The sample contains 8,360 single-class firms with a total of 37,503
firm-year observations and 614 dual-class firms with a total of 2,694
firm-year observations for the period 1995-2002. This table reports
pairwise correlations among the right-hand-side variables used in our
multivariate analysis. Dual is the dual-class status dummy, which
equals one if the firm has multiple share classes, and zero otherwise;
Mktcap is market capitalization, defined as the dollar value of all
share classes at the end of the year (in millions of 2002 dollars);
Return is the annual return on the firm's stock, defined as the
value-weighted average of the returns across traded classes over the
year; Divyield is the dividend yield, defined as the ratio of total
dividend payout to stock price; Retvol is the volatility of stock
returns, defined as the value-weighted average of the stock return
volatility across traded classes using monthly stock returns over the
prior year; Turnover is the share turnover ratio, defined as the
value-weighted average of the ratio of the trading volume to the
number of shares outstanding at the end of the previous year across
all traded classes; M/B is the market-to-book ratio, defined as the
market value of assets divided by the book value of assets; Leverage
is financial leverage, defined as the ratio of total debt to the
market value of assets; Firmage is firm age, defined as the number of
years since the firm first appears in CRSP; Price is the share price,
defined as the value-weighted average of the stock price across traded
classes at the end of the year (in 2002 dollars); S&P500 is an S&P 500
membership dummy, which equals one if the firm is in the S&P 500 index,
and zero otherwise; and #Analysts is analyst coverage, defined as the
number of IBES analysts covering the firm.

 Dual Mktcap Return Divyield

Mktcap -0.008
Return -0.003 0.022 ***
Divyield 0.024 *** 0.060 *** -0.012 **
Retool -0.065 *** -0.090 *** 0.183 * -0.252 ***
Turnover -0.062 *** 0.010 *** 0.065 *** -0.136 ***
M/B -0.040 *** 0.061 *** -0.055 *** -0.125 ***
Leverage 0.067 *** -0.026 *** 0.029 *** 0.191 ***
Firmage 0.018 *** 0.230 *** -0.010 ** 0.387 ***
Price 0.044 *** 0.104 *** 0.005 -0.002
S&P500 -0.009 0.384 *** 0.007 0.155 ***
#Analysts -0.009 0.368 *** -0.016 *** 0.058 ***

 Retool Turnover M/B Leverage

Mktcap
Return
Divyield
Retool
Turnover 0.287 ***
M/B 0.152 *** 0.175 ***
Leverage -0.078 *** -0.155 *** -0.293 ***
Firmage -0.270 *** -0.119 *** -0.130 *** 0.157 ***
Price -0.019 *** -0.007 -0.004 -0.009 *
S&P500 -0.161 *** 0.030 *** 0.026 *** 0.001
#Analysts -0.137 *** 0.186 *** 0.086 *** -0.059 ***

 Firmage Price S&P500

Mktcap
Return
Divyield
Retool
Turnover
M/B
Leverage
Firmage
Price 0.017 ***
S&P500 0.368 *** 0.006
#Analysts 0.237 *** 0.005 *** 0.518 ***

*** Significant at the 0.01 level.

** Significant at the 0.05 level.

* Significant at the 0.10 level.

Table IV. The Effect of Dual-Class Status on Institutional Ownership

The table reports regressions of IO on Dual and lagged control
variables. IO is institutional investors' dollar investment in the
firm's equity as a percentage of the firm's total market value of
equity; Dual is a dummy equal to one if the firm has multiple share
classes, and zero otherwise; Mktcap is the dollar value of all
share classes at the end of the year; Return is the value-weighted
average of the returns across traded classes over the year;
Divyield is the ratio of total dividend payout to stock price;
Retvol is the value-weighted average of the stock return volatility
across traded classes using monthly stock returns over the prior
year; Turnover is the value-weighted average of the ratio of the
trading volume to the number of shares outstanding at the end of
the previous year across all traded classes; M/B is the market
value of assets divided by the book value of assets; Leverage is
the ratio of total debt to the market value of assets; Firmage is
the number of years since the firm first appears in CRSP; Price is
the value-weighted average of the stock price across traded classes
at the end of the year; S&P 500 is a dummy equal to one if the firm
is in the S&P 500 index, and zero otherwise; and #Analysts is the
number of IBES analysts coveting the firm. Panel A reports the
coefficients from pooled OLS regressions with year and 48 Fama and
French industry dummies. The standard errors are clustered at the
firm level. Panel B reports the coefficients on Dual from
Fama-MacBeth regressions with Newey-West standard errors based on
five lags. Panel C reports the coefficients on Dual in a sample
that excludes dual-class firms with both share classes traded. In
all panels, we omit the coefficients on the 48 Fama-French industry
and year dummies. Standard errors are given in parentheses.

 (1) (2) (3) (4)

Panel A. Pooled OLS Regressions with Clustered Standard Errors

Dual -2.551 *** -2.334 *** -3.694 *** -3.597 ***
 (0.871) (0.853) (0.827) (0.826)
Ln(Mktcap) 8.404 *** 7.816 *** 5.158 *** 4.771 ***
 (0.120) (0.133) (0.224) (0.256)
Return -1.026 *** -0.554 *** -1.840 *** -1.772 ***
 (0.134) (0.107) (0.334) (0.325)
Divyield -1.155 *** -1.199 *** -1.599 *** -1.576 ***
 (0.138) (0.139) (0.145) (0.144)
Retvol -6.510 *** -2.406 *** -2.394 ***
 (0.489) (0.506) (0.495)
Turnover 0.139 *** 0.135 *** 0.129 ***
 (0.024) (0.024) (0.024)
M/B -0.660 *** -0.658 ***
 (0.061) (0.061)
Leverage 4.931 *** 4.944 ***
 (0.884) (0.881)
Firmage 0.057 *** 0.055 ***
 (0.020) (0.020)
Ln(Price) 6.213 *** 6.416 ***
 (0.404) (0.403)
S&P500 -0.441
 (1.100)
#Analysts 0.438 **
 (0.183)
Intercept -70.522 *** -61.957 *** -46.949 *** -43.300 ***
 (1.360) (1.524) (1.887) (2.280)
Number of 40,197 40,197 40,197 40,197
 observations
Adjusted 0.446 0.460 0.489 0.490
 [R.sup.2]

Panel B. Fama-MacBeth Regressions with Newey-West Standard Errors.
Includes the Same Controls as in Panel A.

Dual -2.524 *** -2.214 *** -3.681 *** -3.620 ***
 (0.280) (0.351) (0.338) (0.273)
Number of 40,197 40,197 40,197 40,197
 observations

Panel C. Pooled OLS Regressions Using a Subsample Where Dual-Class
Firms with Both Share Classes Traded Are Excluded. Includes the Same
Controls as in Panel A.

Dual -3.636 *** -3.387 *** -4.726 *** -4.651 ***
 (1.038) (1.013) (0.967) (0.965)
Number of 39,451 39,451 39,451 39,451
 observations
Adjusted 0.447 0.461 0.491 0.492
 [R.sup.2]

*** Significant at the 0.01 level.

** Significant at the 0.05 level.

Table V. The Effect of Dual-Class Status on Institutional Ownership by
Type of Institution

The table reports regressions of institutional ownership (IO) by
type of institution on Dual and lagged control variables. IO is
institutional investors' dollar investment in the firm's equity as
a percentage of the firm's total market value of equity; Dual is a
dummy equal to one if the firm has multiple share classes, and zero
otherwise; Mktcap is the dollar value of all share classes at the
end of the year; Return is the value-weighted average of the
returns across traded classes over the year; Divyield is the ratio
of total dividend payout to stock price; Retvol is the
value-weighted average of the stock return volatility across traded
classes using monthly stock returns over the prior year; Turnover
is the value-weighted average of the ratio of the trading volume to
the number of shares outstanding at the end of the previous year
across all traded classes; M/B is the market value of assets
divided by the book value of assets; Leverage is the ratio of total
debt to the market value of assets; Firmage is the number of years
since the firm first appears in CRSP; Price is the value-weighted
average of the stock price across traded classes at the end of the
year (in 2002 dollars); S&P 500 is a dummy equal to one if the firm
is in the S&P 500 index, and zero otherwise; and #Analysts is the
number of IBES analysts covering the firm. Panel A reports the
coefficients from pooled OLS regressions with year and 48
Fama-French industry dummies. The standard errors are clustered at
the firm level. Panel B reports the coefficients on Dual from
Fama-MacBeth regressions with Newey-West standard errors based on
five lags. Panel C reports the coefficients on Dual in a sample
that excludes dual-class firms with both share classes traded. In
all panels, we omit the coefficients on the 48 Fama-French industry
and year dummies. Standard errors are given in parentheses.

 (1) (2) (3)

 Banks Trust Insurance Investment
 Departments Companies Companies

Panel A. Pooled OLS Regressions with Clustered Standard Errors

Dual -0.270 -0.389 *** -0.883 ***
 (0.273) (0.142) (0.298)
Ln(Mktcap) 0.815 *** 0.610 *** 1.287 ***
 (0.053) (0.051) (0.087)
Return -0.261 *** -0.157 *** -0.475 ***
 (0.052) (0.034) (0.092)
Divyield 0.071 * -0.111 *** -0.505 ***
 (0.040) (0.027) (0.047)
Retvol -0.195 *** -0.166 *** -0.650 ***
 (0.065) (0.054) (0.155)
Turnover 0.010 *** 0.007 *** 0.043 ***
 (0.002) (0.002) (0.008)
M/B -0.071 *** -0.039 *** -0.116 ***
 (0.009) (0.006) (0.019)
Leverage 0.295 0.880 *** 0.677 **
 (0.223) (0.231) (0.294)
Firmage 0.043 *** 0.007 * -0.012 *
 (0.004) (0.004) (0.007)
Ln(Price) 0.488 *** 0.196 *** 1.848 ***
 (0.071) (0.064) (0.126)
S&P500 2.318 *** 0.365 * 2.104 ***
 (0.259) (0.188) (0.472)
#Analysts 0.064 ** 0.035 0.528 ***
 (0.032) (0.028) (0.071)
Intercept -7.477 *** -5.656 *** -13.175 ***
 (0.485) (0.466) (0.787)
Number of observations 40,197 40,197 40,197
Adjusted [R.sup.2] 0.351 0.169 0.412

Panel B. Fama-MacBeth Regressions with Newey-West Standard
Errors. Includes the Same Controls as in Panel A.

Dual -0.305 *** -0.381 *** -0.855 ***
 (0.050) (0.034) (0.043)
Number of observations 40,197 40,197 40,197

Panel C. Pooled OLS Regressions Using a Subsample Where Dual-Class
Firms with Both Share Classes Traded Are Excluded. Includes the
Same Controls as in Panel A.

Dual -0.844 *** -0.596 *** -1.231 ***
 (0.194) (0.139) (0.351)
Number of observations 39,451 39,451 39,451
Adjusted [R.sup.2] 0.364 0.170 0.413

 (4) (5)

 Indep. Inv. Other
 Advisors Institutions

Panel A. Pooled OLS Regressions with Clustered Standard Errors

Dual -1.658 *** -0.445 ***
 (0.483) (0.114)
Ln(Mktcap) 1.399 *** 0.662 ***
 (0.137) (0.047)
Return -0.680 *** -0.204 ***
 (0.120) (0.042)
Divyield -0.894 *** -0.146 ***
 (0.080) (0.023)
Retvol -1.447 *** 0.049
 (0.244) (0.072)
Turnover 0.057 *** 0.014 ***
 (0.011) (0.003)
M/B -0.355 *** -0.080 ***
 (0.033) (0.009)
Leverage 3.140 *** 0.021
 (0.537) (0.155)
Firmage 0.011 0.005
 (0.010) (0.003)
Ln(Price) 3.849 *** 0.067
 (0.209) (0.055)
S&P500 -4.699 *** -0.543 ***
 (0.516) (0.180)
#Analysts -0.209 *** 0.014
 (0.078) (0.025)
Intercept -10.517 *** -6.553 ***
 (1.239) (0.442)

Number of observations 40,197 40,197
Adjusted [R.sup.2] 0.295 0.228

Panel B. Fama-MacBeth Regressions with Newey-West Standard
Errors. Includes the Same Controls as in Panel A.

Dual
 -1.633 *** -0.489 **
Number of observations (0.412) (0.149)
 40,197 40,197

Panel C. Pooled OLS Regressions Using a Subsample Where
Dual-Class Firms with Both Share Classes Traded Are Excluded.
Includes the Same Controls as in Panel A.

Dual -1.533 *** -0.470 ***
 (0.578) (0.131)
Number of observations 39,451 39,451
Adjusted [R.sup.2] 0.296 0.228

*** Significant at the 0.01 level.

** Significant at the 0.05 level.

* Significant at the 0.10 level.

Table VI. Changes in Institutional Ownership Following Unification

The table reports regressions of ChgIO, defined as IO in year t + 1
minus IO in year t, where IO is institutional investors' dollar
investment in the firm's equity as a percentage of the firm's total
market value of equity. Unify is a dummy equal to one if the firm
unifies its share classes in year t, and zero otherwise. ChgShouts
is the percentage change in the number of shares outstanding from
year t to year t + 1. Other variables are as of year t. Mktcap is
the dollar value of all share classes at the end of the year;
Return is the value-weighted average of the returns across traded
classes over the year; Divyield is the ratio of total dividend
payout to stock price; Retvol is the value-weighted average of the
stock return volatility across traded classes using monthly stock
returns over the prior year; Turnover is the value-weighted average
of the ratio of the trading volume to the number of shares
outstanding at the end of the previous year across all traded
classes; M/B is the market value of assets divided by the book
value of assets; Leverage is the ratio of total debt to the market
value of assets; Firmage is the number of years since the firm
first appears in CRSP; Price is the value-weighted average of the
stock price across traded classes at the end of the year (in 2002
dollars); S&P 500 is a dummy equal to one if the firm is in the S&P
500 index, and zero otherwise; and #Analysts is the number of IBES
analysts covering the firm. The [DELTA] operator denotes the change
from t - 1 to t. Panel A reports the coefficients from regressions
of ChgIO on Unify and the control variables in levels. Panel B
reports the coefficients from regressions of ChgIO on Unify and the
control variables in changes (except Return, Firmage, and S&P 500).
Panel C reports the coefficients on Unify using a sample of
dual-class firms with only the ordinary classes traded. The control
variables are in levels. Panel D repeats the analysis in Panel C
but uses control variables in changes (except Return, Firmage, and
S&P 500). All regressions include year and 48 Fama-French industry
dummies (not reported), and use standard errors clustered at the
firm level which are given in parentheses.

 (1) (2) (3)
 Institutional Banks Trust Insurance
 Ownership Departments Companies

Panel A. OLS Regressions with Clustered Standard Errors--Controls in
Levels

Unify 10.793 *** 1.393 *** 0.982 **
 (2.141) (0.497) (0.409)
ChgShouts 0.913 -0.015 0.800 ***
 (0.971) (0.247) (0.240)
Ln(Mktcap) 0.338 -0.003 -0.089
 (0.211) (0.054) (0.063)
Return 1.959 *** 0.201 *** 0.208 ***
Divyield (0.093) (0.008) (0.046)
 (0.175) (0.029) (0.043)
Retvol -2.233 *** -0.399 ** -0.191
 (0.855) (0.172) (0.141)
Turnover 0.061 0.016 *** 0.001
 (0.043) (0.006) (0.007)
M/B 0.051 0.039 0.025
 (0.231) (0.063) (0.054)
Leverage -0.248 -0.029 -0.148
 (0.921) (0.209) (0.176)
Firmage -0.032 * 0.003 0.001
 (0.019) (0.003) (0.006)
Ln(Price) 0.883 ** 0.070 0.108 **
 (0.408) (0.076) (0.052)
S&P500 -1.204 -0.08 0.025
 (0.849) (0.185) (0.180)
#Analysts -0.172 0.017 0.023
 (0.194) (0.037) (0.038)
Intercept -6.285 *** -0.046 0.119
 (2.164) (0.571) (0.649)
Number of 2,160 2,160 2,160
observations
Adjusted 0.133 0.039 0.024
[R.sup.2]

Panel B. OLS Regressions with Clustered Standard Errors--Controls
in Changes. The control variables include ChgShouts,
[DELTA]Ln(Mktcap), Return, [DELTA]Divyield, [DELTA]Retvol,
[DELTA]Turnover [DELTA]M/B, [DELTA]Leverage, Firmage,
[DELTA]Ln(Price), S&P500, [DELTA]#Analysts, year dummies, and
industry dummies.

Unify 10.493 *** 1.399 *** 0.952 **
 (2.065) (0.492) (0.411)
Number of 2,160 2,160 2,160
observations
Adjusted 0.162 0.044 0.028
[R.sup.2]

Panel C. OLS Regressions Using a Subsample with Only Ordinary
Shares Traded--Controls in Levels. The control variables include
ChgShouts, Ln(Mktcap), Return, Divyield, Retvol, Turnover, M/B,
Leverage, Firmage, Ln(Price), S&P500, #Analysts, year dummies, and
industry dummies.

Unify 16.850 *** 1.669 ** 1.813 ***
 (4.055) (0.704) (0.592)
Number of 1,552 1,552 1,552
observations
Adjusted 0.159 0.059 0.026
[R.sup.2]

Panel D. OLS Regressions Using a Subsample with Only Ordinary Shares
Traded--Controls in Changes. The control variables include ChgShouts,
[DELTA]Ln(Mktcap), Return, [DELTA]Divyield, [DELTA]Retvol,
[DELTA]Turnover [DELTA]M/B, [DELTA]Leverage, Firmage,
[DELTA]Ln(Price), S&P500, [DELTA]#Analysts, year dummies, and
industry dummies.

Unify 16.077 *** 1.567 ** 1.737 ***
 (3.981) (0.734) (0.583)
Number of 1,552 1,552 1,552
observations
Adjusted 0.191 0.066 0.027
[R.sup.2]

 (4) (5) (6)
 Investment Indep. Inv. Other
 Companies Advisors Institutions

Panel A. OLS Regressions with Clustered Standard Errors--Controls in
Levels

Unify 3.625 *** 3.871 *** 1.020 **
 (0.796) (1.088) (0.419)
ChgShouts -0.151 0.512 -0.200
 (0.511) (0.557) (0.296)
Ln(Mktcap) 0.093 0.246 * 0.114 **
 (0.099) (0.129) (0.047)
Return 0.851 *** 0.667 *** 0.037
Divyield (0.064) (0.012) (0.033)
 (0.091) (0.077) (0.022)
Retvol -1.312 *** -0.509 0.166
 (0.386) (0.394) (0.323)
Turnover 0.005 0.014 0.028 ***
 (0.019) (0.018) (0.009)
M/B 0.214 ** -0.164 -0.077 *
 (0.104) (0.102) (0.046)
Leverage -0.730 * 0.385 0.251
 (0.407) (0.520) (0.239)
Firmage -0.019 ** -0.020 ** 0.005
 (0.008) (0.010) (0.003)
Ln(Price) 0.274 0.296 0.105
 (0.166) (0.199) (0.090)
S&P500 0.698 -1.646 *** -0.263
 (0.454) (0.435) (0.176)
#Analysts -0.147 -0.021 -0.049
 (0.105) (0.096) (0.033)
Intercept -0.929 -5.035 *** -0.609
 (1.002) (1.394) (0.527)
Number of 2,160 2,160 2,160
observations
Adjusted 0.095 0.084 0.082
[R.sup.2]

Panel B. OLS Regressions with Clustered Standard Errors--Controls
in Changes. The control variables include ChgShouts,
[DELTA]Ln(Mktcap), Return, [DELTA]Divyield, [DELTA]Retvol,
[DELTA]Turnover [DELTA]M/B, [DELTA]Leverage, Firmage,
[DELTA]Ln(Price), S&P500, [DELTA]#Analysts, year dummies, and
industry dummies.

Unify 3.454 *** 3.737 *** 1.059 **
 (0.748) (1.058) (0.429)
Number of 2,160 2,160 2,160
observations
Adjusted 0.111 0.095 0.070
[R.sup.2]

Panel C. OLS Regressions Using a Subsample with Only Ordinary
Shares Traded--Controls in Levels. The control variables include
ChgShouts, Ln(Mktcap), Return, Divyield, Retvol, Turnover, M/B,
Leverage, Firmage, Ln(Price), S&P500, #Analysts, year dummies, and
industry dummies.

Unify 4.560 *** 7.358 *** 1.433 **
 (1.459) (2.012) (0.659)
Number of 1,552 1,552 1,552
observations
Adjusted 0.118 0.104 0.09
[R.sup.2]

Panel D. OLS Regressions Using a Subsample with Only Ordinary Shares
Traded--Controls in Changes. The control variables include ChgShouts,
[DELTA]Ln(Mktcap), Return, [DELTA]Divyield, [DELTA]Retvol,
[DELTA]Turnover [DELTA]M/B, [DELTA]Leverage, Firmage,
[DELTA]Ln(Price), S&P500, [DELTA]#Analysts, year dummies, and
industry dummies.

Unify 4.398*** 7.101 *** 1.283 *
 (1.365) (2.010) (0.667)
Number of 1,552 1,552 1,552
observations
Adjusted 0.132 0.113 0.081
[R.sup.2]

*** Significant at the 0.01 level.

** Significant at the 0.05 level.

* Significant at the 0.10 level.

Table VII. Changes in the Number of Shares Held by Institutional
Investors Following Unification

The table reports regressions of %ChgIIShrs, defined as the
percentage change in the number of shares held by institutional
investors from year t to year t + 1. Unify is a dummy equal to one
if the firm unifies its share classes in year t, and zero
otherwise. ChgShouts is the percentage change in the number of
shares outstanding from year t to year t + 1. Other variables are
as of year t. Mktcap is the dollar value of all share classes at
the end of the year; Return is the value-weighted average of the
returns across traded classes over the year; Divyield is the ratio
of total dividend payout to stock price; Retvol is the
value-weighted average of the stock return volatility across traded
classes using monthly stock returns over the prior year; Turnover
is the value-weighted average of the ratio of the trading volume to
the number of shares outstanding at the end of the previous year
across all traded classes; M/B is the market value of assets
divided by the book value of assets; Leverage is the ratio of total
debt to the market value of assets; Firmage is the number of years
since the firm first appears in CRSP; Price is the value-weighted
average of the stock price across traded classes at the end of the
year (in 2002 dollars); S&P500 is a dummy equal to one if the firm
is in the S&P 500 index, and zero otherwise; and #Analysts is the
number of IBES analysts coveting the firm. The [DELTA] operator
denotes the change from t-1 to t. Panel A reports the coefficients
from regressions of %ChgIIShrs on Unify and the control variables
in levels. Panel B reports the coefficients from regressions of
%ChgIIShrs on Unify and the control variables in changes (except
Return, Firmage, and S&P500). Panel C reports the coefficients on
Unify using a sample of dual-class firms with only the ordinary
classes traded. The control variables are in levels. Panel D
repeats the analysis in Panel C but uses control variables in
changes (except Return, Firmage, and S&P500). All regressions
include year and 48 Fama-French industry dummies (not reported),
and use standard errors clustered at the firm level, which are
given in parentheses.

 (1) (2) (3)
 Institutional Banks Trust Insurance
 Ownership Departments Companies

Panel A. OLS Regressions with Clustered Standard Errors--Controls
in Levels. The control variables include ChgShouts, Ln(Mktcap),
Return, Divyield, Retvol, Turnover, M/B, Leverage, Firmage,
Ln(Price), S&P500, #Analysts, year dummies, and industry
dummies.

 0.335 *** 0.424 *** 0.287
 (0.090) (0.145) (0.182)
Number of 2,121 1,994 1,681
 observations
Adjusted 0.324 0.177 0.141
 [R.sup.2]

Panel B. OLS Regressions with Clustered Standard Errors--Controls
in Changes. The control variables include ChgShouts,
[DELTA]Ln(Mktcap), Return, [DELTA]Divyield, [DELTA]Retvol,
[DELTA]Turnover. [DELTA]M/B, [DELTA]Leverage, Firmage,
[DELTA]Ln(Price), S&P500, [DELTA]#Analysts, year dummies, and
industry dummies.

Unify 0.316 *** 0.384 *** 0.276
 (0.084) (0.135) (0.182)
Number of 2,121 1,994 1,681
 observations
Adjusted 0.341 0.209 0.149
 [R.sup.2]

Panel C. OLS Regressions Using a Subsample with Only Ordinary
Shares Traded--Controls in Levels. The control variables include
ChgShouts, Ln(Mktcap), Return, Divyield, Retvol, Turnover, M/B,
Leverage, Firmage, Ln(Price), S&P500, #Analysts, year dummies, and
industry dummies.

Number of 1,528 1,426 1,215
 observations
Adjusted 0.354 0.202 0.142
 [R.sup.2]

Panel D. OLS Regressions Using a Subsample with Only Ordinary
Shares Traded-Controls in Changes. The control variables include
ChgShouts, OLn(Mktcap), Return, ODivyield, ORetvol, A Turnover
OM/B, ALeverage, Firmage, ALn(Price), S&P500, A#Analysts, year
dummies, and industry dummies.

Unify 0.583 *** 0.524 ** 0.826 ***
 (0.151) (0.203) (0.315)
Number of 1,528 1,426 1,215
 observations
Adjusted 0.354 0.224 0.154
 [R.sup.2]

 (4) (5) (6)
 Investment Indep. Inv. Other
 Companies Advisors Institutions

Panel A. OLS Regressions with Clustered Standard Errors--Controls
in Levels. The control variables include ChgShouts, Ln(Mktcap),
Return, Divyield, Retvol, Turnover, M/B, Leverage, Firmage,
Ln(Price), S&P500, #Analysts, year dummies, and industry
dummies.

 0.385 *** 0.198 ** 0.295 *
 (0.123) (0.096) (0.159)
Number of 1,844 2,079 1,643
 observations
Adjusted 0.194 0.271 0.151
 [R.sup.2]

Panel B. OLS Regressions with Clustered Standard Errors--Controls
in Changes. The control variables include ChgShouts,
[DELTA]Ln(Mktcap), Return, [DELTA]Divyield, [DELTA]Retvol,
[DELTA]Turnover. [DELTA]M/B, [DELTA]Leverage, Firmage,
[DELTA]Ln(Price), S&P500, [DELTA]#Analysts, year dummies, and
industry dummies.

Unify 0.371 *** 0.186 ** 0.267 *
 (0.121) (0.094) (0.153)
Number of 1,844 2,079 1,643
 observations
Adjusted 0.206 0.275 0.170
 [R.sup.2]

Panel C. OLS Regressions Using a Subsample with Only Ordinary
Shares Traded--Controls in Levels. The control variables include
ChgShouts, Ln(Mktcap), Return, Divyield, Retvol, Turnover, M/B,
Leverage, Firmage, Ln(Price), S&P500, #Analysts, year dummies, and
industry dummies.

Number of 1,339 1,497 1,201
 observations
Adjusted 0.218 0.295 0.177
 [R.sup.2]

Panel D. OLS Regressions Using a Subsample with Only Ordinary
Shares Traded-Controls in Changes. The control variables include
ChgShouts, OLn(Mktcap), Return, ODivyield, ORetvol, A Turnover
OM/B, ALeverage, Firmage, ALn(Price), S&P500, A#Analysts, year
dummies, and industry dummies.

Unify 0.566 *** 0.413 ** 0.577 ***
 (0.218) (0.172) (0.218)
Number of 1,339 1,497 1,201
 observations
Adjusted 0.237 0.301 0.202
 [R.sup.2]

*** Significant at the 0.01 level.

** Significant at the 0.05 level.

* Significant at the 0.10 level.

Table VIII. Institutional Ownership and Future Unification Decisions

The table reports marginal effects of probit models of the
probability that a dual-class firm unifies its share classes. The
dependent variable, Unify, equals one if the firm abandons its
dual-class structure in year t, and zero otherwise. IO is
institutional investors' dollar investment in the firm's equity as
a percentage of the firm's total market value of equity. Similarly,
BTIO is the ownership in the firm by the bank's trust departments;
INSCIO is the ownership by insurance companies; INVCIO is the
ownership by investment companies; IAIO is the ownership by
independent advisors; and OIIO is the ownership by the other types
of institutions. #Mergers is the number of mergers in the firm's
Fama-French industry; Capex is capital expenditures divided by
assets; ROA is the firm's return on assets; Mktcap is the dollar
value of all share classes at the end of the year; Return is the
value-weighted average of the returns across traded classes over
the year; Divyield is the ratio of total dividend payout to stock
price; Retvol is the value-weighted average of the stock return
volatility across traded classes using monthly stock returns over
the prior year; Turnover is the value-weighted average of the ratio
of the trading volume to the number of shares outstanding at the
end of the previous year across all traded classes; M/B is the
market value of assets divided by the book value of assets;
Leverage is the ratio of total debt to the market value of assets;
Firmage is the number of years since the firm first appears in
CRSP; Price is the value-weighted average of the stock price across
traded classes at the end of the year (in 2002 dollars); S&P500 is
a dummy equal to one if the firm is in the S&P 500 index, and zero
otherwise; #Analysts is the number of IBES analysts covering the
firm; Distress is a dummy equal to one if the firm's cash flow is
less than its interest payments, and zero otherwise; Retearn is the
ratio of retained earnings to total assets; and #Shares is the
total number of shares outstanding across all share classes. Let
P(x) be the probability that a firm unifies its dual-class
structure in year t, and let [DELTA] be the change of a variable
between year t and year t-1. Panel A estimates P(x) as a function
of lagged institutional ownership and control variables:

[MATHEMATICAL EXPRESSION NOT REPRODUCIBLE IN ASCII]

Panel B estimates P(.) as a function of lagged institutional
ownership and contemporaneous changes in the control variables:

[MATHEMATICAL EXPRESSION NOT REPRODUCIBLE IN ASCII]

Panel C estimates P(-) as a function of lagged changes
institutional ownership and contemporaneous changes in the control
variables:

[MATHEMATICAL EXPRESSION NOT REPRODUCIBLE IN ASCII]

The coefficients on the institutional ownership variables are
multiplied by 1,000. All regressions include year and 48
Fama-French industry dummies and adjust for the clustering of
observations at the firm level. The coefficients on the control
variables, industry dummies, and year dummies are omitted for
brevity. Standard errors are given in parentheses.

 (1) (2) (3) (4)

Panel A. Probability of Unification as a Function of Lagged
Institutional Ownership and Control Variables. The control
variables include [#Mergers.sub.t-1], [Capex.sub.t-1],
[Ret.sub.t-1], [ROA.sub.t-1], [Ln(Mktcap) .sub.t-1],
[Divyield.sub.t-1], [Retvol.sub.t-1], [Turnover.sub.t-1],
[M/B.sub.t-1], [Leverage.sub.t-1], [Firmage.sub.t-1], [Ln(Price)
.sub.t-1], [S&P500.sub.t-1], [#Analysts.sub.t-1],
[Distress.sub.t-1], [Retearn.sub.t-1], industry dummies, and year
dummies.

[I0.sub.t-1] x 1000 -0.022
 (0.186)
[BTIO.sub.t-1] x 1000 -0.149
 (0.664)
[INSCIO.sub.t-1] x 1000 0.418
 (0.540)
[INVCIO.sub.t-1] x 1000 -0.019
 (0.462)
[IAIO.sub.t-1] x 1000

[OIIO.sub.t-1] x 1000

Panel B. Probability of Unification as a Function of Lagged
Institutional Ownership and Contemporaneous Changes in the Control
Variables. The control variables include [#Mergers.sub.t-1],
[DELTA][Capex.sub.t-1], [Ret.sub.t-1], [ROA.sub.t-1],
[DELTA]Ln(Mktcap), [DELTA]Divyield, [DELTA]Retvol, [DELTA]Turnover,
[DELTA]M/B, [DELTA]Leverage, [DELTA]Firmage, [DELTA]Ln(Price),
[S&P500.sub.t-1], [DELTA]#Analysts, [Distresst.sub.t-1],
[DELTA]Retearn, [DELTA]#Shares, industry dummies, and year dummies.

[I0.sub.t-1] x 1000 0.056
 (0.154)
[BTIO.sub.t-1] x 1000 -0.045
 (0.468)
[INSCIO.sub.t-1] x 1000 0.550
 (0.544)
[INVCIO.sub.t-1] x 1000 -0.072
 (0.382)
[IAIO.sub.t-1] x 1000

[OIIO.sub.t-1] x 1000

Number of observations 1,233 1,233 1,233 1,233

Panel C. Probability of Unification as a Function of Lagged Change
in Institutional Ownership and Contemporaneous Changes in the
Control Variables. The control variables include
[#Mergers.sub.t-1], [DELTA]Capex, [Ret.sub.t-1], [ROA.sub.t-1],
[DELTA]Ln(Mktcap), [DELTA]Divyield, [DELTA]Retvol, [DELTA]Turnover,
[DELTA]M/B, [DELTA]Leverage, [DELTA]Firmage, [DELTA]Ln(Price),
[S&P500.sub.t-1], [DELTA]#Analysts, [Distress.sub.t-1],
[DELTA]Retearn, [DELTA]#Shares, industry dummies, and year dummies.

[DELTA][I0.sub.t-1] x 1000 -0.392
 (0.260)
[I0.sub.t-2] x 1000 0.110
 (0.174)
[DELTA][BTIO.sub.t-1] x 1000 -1.222
 (1.173)
[BTIO.sub.t-2] x 1000 -0.079
 (0.528)
[DELTA][INSCIO.sub.t-1] x 1000 0.517
 (0.452)
[INSCIO.sub.t-2] x 1000 1.050
 (0.954)
[DELTA][INVCIO.sub.t-1] x 1000 -0.860
 (0.699)
[INVCIO.sub.t-2] x 1000 -0.244
 (0.449)
[DELTA][IAIO.sub.t-1] x 1000

[IAIO.sub.t-2] x 1000

[DELTA][OIIO.sub.t-1] x 1000

[OIIO.sub.t-2] x 1000

Number of observations 1,600 1,600 1,600 1,600
Number of observations 1,588 1,588 1,588 1,588

 (5) (6) (7)

Panel A. Probability of Unification as a Function of Lagged
Institutional Ownership and Control Variables. The control
variables include [#Mergers.sub.t-1], [Capex.sub.t-1],
[Ret.sub.t-1], [ROA.sub.t-1], [Ln(Mktcap) .sub.t-1],
[Divyield.sub.t-1], [Retvol.sub.t-1], [Turnover.sub.t-1],
[M/B.sub.t-1], [Leverage.sub.t-1], [Firmage.sub.t-1], [Ln(Price)
.sub.t-1], [S&P500.sub.t-1], [#Analysts.sub.t-1],
[Distress.sub.t-1], [Retearn.sub.t-1], industry dummies, and year
dummies.

[I0.sub.t-1] x 1000

[BTIO.sub.t-1] x 1000 -0.086
 (0.624)
[INSCIO.sub.t-1] x 1000 0.441
 (0.550)
[INVCIO.sub.t-1] x 1000 0.033
 (0.460)
[IAIO.sub.t-1] x 1000 -0.088 -0.066
 (0.296) (0.294)
[OIIO.sub.t-1] x 1000 -0.622 -0.582
 (1.214) (1.232)

Panel B. Probability of Unification as a Function of Lagged
Institutional Ownership and Contemporaneous Changes in the Control
Variables. The control variables include [#Mergers.sub.t-1],
[DELTA][Capex.sub.t-1], [Ret.sub.t-1], [ROA.sub.t-1],
[DELTA]Ln(Mktcap), [DELTA]Divyield, [DELTA]Retvol, [DELTA]Turnover,
[DELTA]M/B, [DELTA]Leverage, [DELTA]Firmage, [DELTA]Ln(Price),
[S&P500.sub.t-1], [DELTA]#Analysts, [Distresst.sub.t-1],
[DELTA]Retearn, [DELTA]#Shares, industry dummies, and year dummies.

[I0.sub.t-1] x 1000

[BTIO.sub.t-1] x 1000 -0.097
 (0.475)
[INSCIO.sub.t-1] x 1000 0.563
 (0.558)
[INVCIO.sub.t-1] x 1000 -0.211
 (0.387)
[IAIO.sub.t-1] x 1000 0.100 0.120
 (0.274) (0.266)
[OIIO.sub.t-1] x 1000 0.167 0.110
 (1.082) (1.134)
Number of observations 1,233 1,233 1,233

Panel C. Probability of Unification as a Function ofLagged Change
in Institutional Ownership and Contemporaneous Changes in the
Control Variables. The control variables include
[#Mergers.sub.t-1], [DELTA]Capex, [Ret.sub.t-1], [ROA.sub.t-1],
[DELTA]Ln(Mktcap), [DELTA]Divyield, [DELTA]Retvol, [DELTA]Turnover,
[DELTA]M/B, [DELTA]Leverage, [DELTA]Firmage, [DELTA]Ln(Price),
[S&P500.sub.t-1], [DELTA]#Analysts, [Distress.sub.t-1],
[DELTA]Retearn, [DELTA]#Shares, industry dummies, and year dummies.

[DELTA][I0.sub.t-1] x 1000

[I0.sub.t-2] x 1000

[DELTA][BTIO.sub.t-1] x 1000 -0.290
 (0.960)
[BTIO.sub.t-2] x 1000 -0.086
 (0.455)
[DELTA][INSCIO.sub.t-1] x 1000 0.233
 (0.538)
[INSCIO.sub.t-2] x 1000 0.692
 (0.985)
[DELTA][INVCIO.sub.t-1] x 1000 -0.569
 (0.577)
[INVCIO.sub.t-2] x 1000 -0.417
 (0.463)
[DELTA][IAIO.sub.t-1] x 1000 -0.669 -0.450
 (0.458) (0.443)
[IAIO.sub.t-2] x 1000 0.193 0.239
 (0.305) (0.313)
[DELTA][OIIO.sub.t-1] x 1000 -1.423 -0.795
 (1.264) (1.167)
[OIIO.sub.t-2] x 1000 0.696 0.701
 (1.405) (1.392)
Number of observations 1,600 1,600 1,600
Number of observations 1,588 1,588 1,588

Table IX. Controlling for Insider Ownership and Board Characteristics

The table reports OLS regressions of 10 on Dual and lagged control
variables. 10 is institutional investors' dollar investment in the
firm's equity as a percentage of the firm's total market value of
equity. Dual is a dummy equal to one if the firm has multiple share
classes, and zero otherwise. InsiderOwn is the percentage of the
firm's equity held by insiders. BoardSize is the number of
directors serving on the board. Pctindep is the percentage of the
board members that are independent directors. CEOCOB is a dummy
equal to one if the CEO is also the chairman of the board, and zero
otherwise. All other control variables are the same as in Table IV,
and are omitted for brevity. Mktcap is the dollar value of all
share classes at the end of the year; Return is the value-weighted
average of the returns across traded classes over the year;
Divyield is the ratio of total dividend payout to stock price;
Retvol is the value-weighted average of the stock return volatility
across traded classes using monthly stock returns over the prior
year; Turnover is the value-weighted average of the ratio of the
trading volume to the number of shares outstanding at the end of
the previous year across all traded classes; MIB is the market
value of assets divided by the book value of assets; Leverage is
the ratio of total debt to the market value of assets; Firmage is
the number of years since the firm first appears in CRSP; Price is
the value-weighted average of the stock price across traded classes
at the end of the year (in 2002 dollars); S&PS00 is a dummy equal
to one if the firm is in the S&P 500 index, and zero otherwise;
#Analysts is the number of IBES analysts covering the firm. All
regressions include year and 48 Fama-French industry dummies, and
adjust for the clustering of observations at the firm level.
Standard errors are given in parentheses.

 (1) (2) (3)

Dual -9.604 *** -8.836 *** -9.414 ***
 (1.411) (1.339) (1.421)
InsiderOwn -0.639 ***
 (0.036)
BoardSize -0.627 ***
 (0.146)
PctIndep

CEOCOB

Number of 8,079 8,079 8,079
 observations
Adjusted 0.336 0.408 0.341
 [R.sup.2]

 (4) (5) (6)

Dual -7.542 *** -9.506 *** -7.578 ***
 (1.473) (1.418) (1.401)
InsiderOwn -0.609 ***
 (0.036)
BoardSize -0.796 ***
 (0.134)
PctIndep 0.171 *** 0.079***
 (0.019) (0.018)
CEOCOB 1.049 1.067 *
 (0.648) (0.597)
Number of 8,079 8,079 8,079
 observations
Adjusted 0.359 0.337 0.422
 [R.sup.2]

*** Significant at the 0.01 level.

* Significant at the 0.10 level.

Table X. Controlling for Institutional Herding

The table reports regressions of IO by type of institution on Dual
and lagged control variables. IO is institutional investors' dollar
investment in the firm's equity as a percentage of the firm's total
market value of equity; the Lakonishok, Shleifer, and Vishny (1992)
proxy for institutional investors' herding in their investment
decisions, Herd, is the number of institutions that are net buyers
of the firm's stock in a given year divided by the sum of the
number of institutions that are net buyers and those that are net
sellers of the firm's stock in the same year; Dual is a dummy equal
to one if the firm has multiple share classes, and zero otherwise;
Mktcap is the dollar value of all share classes at the end of the
year; Return is the value-weighted average of the returns across
traded classes over the year; Divyield is the ratio of total
dividend payout to stock price; Retvol is the value-weighted
average of the stock return volatility across traded classes using
monthly stock returns over the prior year; Turnover is the
value-weighted average of the ratio of the trading volume to the
number of shares outstanding at the end of the previous year across
all traded classes; M/B is the market value of assets divided by
the book value of assets; Leverage is the ratio of total debt to
the market value of assets; Firmage is the number of years since
the firm first appears in CRSP; Price is the value-weighted average
of the stock price across traded classes at the end of the year;
S&P500 is a dummy equal to one if the firm is in the S&P 500 index,
and zero otherwise; and #Analysts is the number of IBES analysts
covering the firm. All regressions include year and 48 Fama-French
industry and year dummies (not reported), and adjust for the
clustering of observations at the firm level. Standard errors are
given in parentheses.

 (1) (2) (3)
 Institutional Banks Trust Insurance
 Ownership Departments Companies

Dual -4.207 *** -0.31 -0.406 ***
 -0.812 -0.273 -0.142
Herd 11.194 *** 0.724 *** 0.303 ***
 -0.411 -0.078 -0.066
Number of 40,197 40,197 40,197
 observations
Adjusted 0.505 0.353 0.169
 [R.sup.2]

 (4) (5) (6)
 Investment Indep. Inv. Other
 Companies Advisors Institutions

Dual -0.987 *** -2.059 *** -0.492 ***
 -0.296 -0.473 -0.113
Herd 1.911 *** 7.363 *** 0.862 ***
 -0.152 -0.231 -0.074
Number of 40,197 40,197 40,197
 observations
Adjusted 0.415 0.321 0.231
 [R.sup.2]

*** Significant at the 0.01 level.

Table XI. Different Assumptions for the Voting Premium

The table reports OLS regressions of aggregate institutional
ownership and by type of institution on Dual and lagged control
variables. Dual is a dummy equal to one if the firm has multiple
share classes, and zero otherwise; Mktcap is the dollar value of
all share classes at the end of the year; Return is the
value-weighted average of the returns across traded classes over
the year; Divyield is the ratio of total dividend payout to stock
price; Retvol is the value-weighted average of the stock return
volatility across traded classes using monthly stock returns over
the prior year; Turnover is the value-weighted average of the ratio
of the trading volume to the number of shares outstanding at the
end of the previous year across all traded classes; M/B is the
market value of assets divided by the book value of assets;
Leverage is the ratio of total debt to the market value of assets;
Firmage is the number of years since the firm first appears in
CRSP; Price is the value-weighted average of the stock price across
traded classes at the end of the year; S&P500 is a dummy equal to
one if the firm is in the S&P 500 index, and zero otherwise; and
#Analysts is the number of IBES analysts coveting the firm. All
regressions include year and 48 Fama-French industry and year
dummies (not reported) and adjust for the clustering of
observations at the firm level. In Panel A, institutional ownership
is defined as institutional investors' dollar investment in the
firm's equity as a percentage of the firm's total market value of
equity (nontraded superior-voting shares are assumed to have a
price that is 5% higher than the price of traded inferior voting
shares). In Panel B, institutional ownership is defined as
institutional investors' dollar investment in the firm's equity as
a percentage of the firm's total market value of equity (nontraded
superior-voting shares are assumed to have a price that is 10%
higher than the price of traded inferior voting shares). Standard
errors are given in parentheses.

 (1) (2) (3)

 Institutional Banks Trust Insurance
 Ownership Departments Companies

Panel A. Using 5% Voting Premium to Compute IO. Includes all the
controls in Tables IV and V.

Dual -3.992 *** -0.317 -0.416 ***
 (0.828) (0.273) (0.142)
Number of 40,197 40,197 40,197
 observations
Adjusted 0.489 0.351 0.169
 [R.sup.2]

Panel B. Using 10% Voting Premium to Compute IO. Includes all the
controls in Tables IV and V.

Dual -4.379 *** -0.362 -0.443 ***
 (0.831) (0.273) (0.141)
Number of 40,197 40,197 40,197
 observations
Adjusted 0.489 0.351 0.169
 [R.sup.2]

 (4) (5) (6)

 Investment Indep. Inv. Other
 Companies Advisors Institutions

Panel A. Using 5% Voting Premium to Compute IO. Includes all the
controls in Tables IV and V.

Dual -0.983 *** -1.855 *** -0.472 ***
 (0.298) (0.483) (0.114)
Number of 40,197 40,197 40,197
 observations
Adjusted 0.412 0.295 0.228
 [R.sup.2]

Panel B. Using 10% Voting Premium to Compute IO. Includes all the
controls in Tables IV and V.

Dual -1.080 *** -2.047 *** -0.499 ***
 (0.297) (0.482) (0.113)
Number of 40,197 40,197 40,197
 observations
Adjusted 0.412 0.294 0.228
 [R.sup.2]

*** Significant at the 0.01 level.

Table XII. Matched Sample Approach

The sample contains 2,694 dual-class firm-year observations and an
equal number of single-class firm-year observations for the period
1995-2002. The single-class control group is obtained by matching
the dual-class firms in our sample based on 48 Fama-French industry
and total assets. The table reports regressions of aggregate
institutional ownership and by type of institution on Dual and
lagged control variables. IO is institutional investors' dollar
investment in the firm's equity as a percentage of the firm's total
market value of equity; Dual is a dummy equal to one if the firm
has multiple share classes, and zero otherwise; Mktcap is the
dollar value of all share classes at the end of the year; Return is
the value-weighted average of the returns across traded classes
over the year; Divyield is the ratio of total dividend payout to
stock price; Retvol is the value-weighted average of the stock
return volatility across traded classes using monthly stock returns
over the prior year; Turnover is the value-weighted average of the
ratio of the trading volume to the number of shares outstanding at
the end of the previous year across all traded classes; M/B is the
market value of assets divided by the book value of assets;
Leverage is the ratio of total debt to the market value of assets;
Firmage is the number of years since the firm first appears in
CRSP; Price is the value-weighted average of the stock price across
traded classes at the end of the year; S&P500 is a dummy equal to
one if the firm is in the S&P 500 index, and zero otherwise; and
#Analysts is the number of IBES analysts covering the firm. The
control variables are omitted for brevity. Panel A reports the
coefficients on Dual from pooled OLS regressions which include year
dummies and 48 Fama-French industry dummies. The standard errors
are adjusted for the clustering of observations at the firm level.
Panel B reports the coefficients on Dual from Fama-MacBeth
regressions with Newey-West standard errors based on five lags.
Standard errors are given in parentheses.

 (1) (2) (3)
 Institutional Banks Trust Insurance
 Ownership Departments Companies

Panel A. Pooled OLS Regressions with Clustered Standard Errors.
Includes all the controls in Tables IV and V.

Dual -3.295 *** -0.199 -0.464 **
 -0.979 -0.294 -0.186
Number of 5,388 5,388 5,388
 observations
Adjusted 0.423 0.251 0.134
 [R.sup.2]

Panel B. Fama-MacBeth Regressions with Newey-West Standard Errors.
Includes all the controls in Tables IV and V.

Dual -3.295 *** -0.220 -0.486 ***
 -0.622 -0.125 -0.032
Number of 5,388 5,388 5,388
 observations

 (4) (5) (6)
 Investment Indep. Inv. Other
 Companies Advisors Institutions

Panel A. Pooled OLS Regressions with Clustered Standard Errors.
Includes all the controls in Tables IV and V.

Dual -0.453 -1.853 *** -0.352 **
 -0.340 -0.586 -0.139
Number of 5,388 5,388 5,388
 observations
Adjusted 0.349 0.269 0.285
 [R.sup.2]

Panel B. Fama-MacBeth Regressions with Newey-West Standard Errors.
Includes all the controls in Tables IV and V.

Dual -0.408 *** -1.768 ** -0.430 **
 -0.094 -0.588 -0.155
Number of 5,388 5,388 5,388
 observations

*** Significant at the 0.01 level.

** Significant at the 0.05 level.
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Author:Li, Kai; Ortiz-Molina, Hernan; Zhao, Xinlei
Publication:Financial Management
Article Type:Report
Geographic Code:1CANA
Date:Dec 22, 2008
Words:19393
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