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The Organizational Justice-Job Engagement Relationship: How Social Exchange and Identity Explain This Effect.

Effective leadership involves motivating employees to contribute to firm goals (Yukl, 2013). Organizational leaders motivate employees to contribute to firm goals by creating environments that engage their employees organization-wide (Gallup Management Journal, 2013). An engaged workforce, one in which employees are motivated to invest physical, cognitive, and emotional energies into work roles (i.e., job engagement; Macey and Schneider, 2008), serves as a primary means of competitive advantage and overall organizational effectiveness (e.g., Barrick et al., 2015). Given the substantial influence organizational leaders have over the rewards, policies, and procedures guiding employee work behaviors, leaders have the ability to impact the job engagement process for all employees. A central way to do so is by offering equitable outcomes (i.e., distributive justice) and enacting fair policies and procedures (i.e., procedural justice) (Colquitt et al, 2005). Indeed, the positive effects of these justice judgments on job engagement are well supported (e.g., Haynie et al, 2016; Saks, 2006). What is still unclear, however, is how organizational justice promotes job engagement.

To better understand this relationship, researchers have advanced two perspectives: the group engagement model and social exchange theory. In the group engagement model, employees become more engaged when they are treated justly at work primarily from a connection with their organization manifested through deeper organizational identity (Tyler and Blader, 2003). Advancing this approach, Tyler and Blader (2000) and Blader and Tyler (2009) demonstrated that organizational identification (i.e., a psychological connection between an employee and his/her organization; Ashforth and Mael, 1989) explained how distributive and procedural justice enhance self-reported behavioral intentions and discretionary work behaviors. Additionally, research has shown organizational justice's capacity for enhancing employee-assessed job engagement (e.g., Ghosh et al, 2014; Sharoni et al, 2015) where a substantial portion of this motivational effect can be attributed to organizational identification (Heetal, 2014; Olkkonen and Lipponen, 2006).

Independently, researchers have also considered a social exchange perspective to explain how justice impacts job engagement. Proponents of social exchange argue that when individuals receive just treatment, they respond in kind by engaging in behaviors that are desirable to the other party (Blau, 1964; Colquitt et al, 2013; Cropanzano and Mitchell, 2005). Distributive and procedural justice clarify behavioral expectations and promote stronger social exchange relationships with the organization. One representation of this reciprocal relationship is perceived organizational support (POS) (i.e., the perception that the organization cares about employees' well-being and values their contributions; Ahmed and Nawaz, 2015; Rhoades and Eisenberger, 2002). These supportive gestures comfort employees by enhancing their sense of psychological security (Tyler and Lind, 1996) so job engagement more easily materializes.

While researchers have used these two contrasting theoretical perspectives to identify mediating effects for organizational identification and perceived organizational support, respectively, they have not yet considered these constructs together. This study addresses this gap by considering a dual-mediated pathway that explains the influence of justice on employee job engagement. The current research contributes to the justice and engagement literatures by first offering a deeper understanding of the extent social exchange relationships stemming from organizational justice impact employee social identities that in turn enhance job engagement. The integration of the social exchange and group engagement model offers a better understanding of employees' engagement and the rationale used when motivated by fair organizational treatment. Second, implications from this research exist for broader identity theories to consider social exchange tenets in their identity models. Social identity is a valuable psychological motivator directing many workplace behaviors (Haslam, 2004), so it is possible that these identities occur in response to normative expectations found within social exchange relationships (Gouldner, 1960), especially when responding to fair treatment.

HYPOTHESES DEVELOPMENT

Justice, Identity, and Job Engagement

Organizational leaders are perceived to be well intentioned when they offer fair outcomes, policies, and procedures to employees (Colquitt et al., 2005). In response, employees develop justice perceptions that guide their work efforts (Lewin, 1936). Because job engagement requires a deeper investment by employees, just treatment signals to employees that the organization has their best interests in mind, providing a sense of comfort and reducing any fears of being taken advantage of (Tyler and Lind, 1996). By removing exploitation concerns, employees can utilize their work energies to become more job engaged (Kahn, 1990).

The group engagement model, in particular, suggests the effects of justice on job engagement stem from a powerful psychological connection, or identity, with the employing organization (Tyler and Blader, 2003). Individuals are intrinsically motivated to assist in the success of groups with which they identify (Tajfel, 1978) and thus are motivated to do what is in the best interest of the group. Within organizations, employees holding salient organizational identities attempt to ensure the viability of their organization by seeking to meet behavioral expectations (Riketta, 2005) such as through heightened levels of job engagement (Christian et al., 2011). Thus, organizational identification may explain how justice impacts job engagement. Tyler and Blader (2003) focused their group engagement model on distributive and procedural justice because when policies and outcomes are consistent and fair, employees feel more identity security so their social identities more easily form. He et al. (2014) found this mediating effect for the procedural justice-engagement relationship but did not account for distributive justice in their model. Thus, it is important to examine the simultaneous positive influence of both distributive and procedural justice on organizational identification and subsequent job engagement to potentially capture nuanced effects unique to each.

Hypothesis 1: Organizational identification mediates the relationships of (a) distributive and (b) procedural justice with job engagement.

The Role of Social Exchange in Group Engagement

While distributive and procedural justice were hypothesized to predict organizational identification and job engagement in turn, employees must internalize the group norms and expectations that guide work efforts for these effects to occur. Work efforts relating to behavioral expectations are thus contingent on the extent norms activated by social identities prescribe these behaviors (Haslam, 2004; Van Knippenberg and Ellemers, 2003). This normative guidance may be attributed to a social exchange relationship formed between employees and the organization.

Distributive and procedural justice represent two mechanisms by which the organization can enhance organizationally-directed social exchange relationships that promote employees' workplace reciprocation desires (e.g., Masterson et al., 2000). Because social exchanges are guided by a norm of reciprocity, employees receiving beneficial treatment such as justice should be more prone to respond in kind with behavioral expectations beneficial to the party extending the desirable treatment (Kelley and Thibault, 1978). The understanding brought on by fair organizational outcomes (distributive justice) and consistent and bias free processes for achieving those outcomes (procedural justice) provide the behavioral expectation clarity so social exchange relationships more easily form.

The quality of the social exchange relationship with the organization can be represented by perceived organizational support (POS; Rhoades and Eisenberger, 2002). POS is a recognition by employees that their organization treats them favorably and provides discretionary resources to them, which motivates the employees to engage in more positive work for the organization. Employees who perceive high distributive and procedural justice are likely to form stronger social exchange relationships with their organization, as indicated by higher ratings of POS (Masterson et al,, 2000; Wayne et al., 2002). Once social exchange expectations are clarified as found with high POS, employees are more likely to integrate the organization into their self-concept, leading to increased organizational identification. Indeed, such a process is likely required in order to alleviate employees' exploitation concerns so they perceive less risk in merging their identity with that of the organization. Therefore, POS has the potential for explaining the justice-organizational identification relationship in much the same way that the favorable risk-to-reward ratio captured in calculative trust must be developed and tested prior to establishing a deeper relational connection of identity-based trust (Rousseau et al., 1998).

Further, as the exchange relationship requires mutual effort from both parties (Blau, 1964), employees who recognize the benefits of POS are indebted to the organization and respond in kind. Such a reciprocal response may be manifested in heightened job engagement (Crawford et al., 2010) and are likely due, at least in part, to a deepening of organizational identification. This combined effect suggests that POS and organizational identification may sequentially explain the relationships of distributive and procedural justice with job engagement.

Hypothesis 2: Perceived organizational support mediates the relationships of (a) distributive and (b) procedural justice with organizational identification.

Hypothesis 3: The relationships of (a) distributive and (b) procedural justice are indirectly related to job engagement first through perceived organizational support and then organizational identification.

METHOD

The theoretical model was tested independently across two samples using the same methodology to improve the generalizability of the findings.

Procedures and Samples

In Sample 1, employees of an engineering firm in the southeastern United States were sampled. Data were collected through a voluntary online survey. Of the 966 employees contacted for participation, 308 completed the survey (31.88% response rate). Employee tenure (in years) was collected from archival sources to test for nonresponse bias between respondents and those who self-selected out of the study. Results of the independent-samples t-test indicated no difference in tenure ([chi square](964) = 0.82, p 0.41). Average tenure of respondents for Sample 1 is 3.24 years (s.d. = 3.07).

Given the potential homogeneity of the first sample, a separate, more diverse sample (Sample 2) was collected using a snowballing technique which allowed for the collection of data from a wide array of industries and job types (Zinkhan et al., 1983). Undergraduate students recruited 3-5 working adults from an organization in which they worked or were associated. If students were employed in the organization, they were permitted to participate as one of the employees. The collection methodology and instruments in this study were identical to those used in the first sample. A total of 460 respondents completed the survey with an average tenure for Sample 2 of 4.56 years (s.d. = 5.68).

Study Measures

All measures were recorded on a Likert scale ranging from 1 (strongly disagree) to 5 (strongly agree) unless otherwise noted. Cronbach's alphas for Samples 1 and 2 are referenced as [[alpha].sub.1] and [[alpha].sub.2], respectively.

OrganizationalJustice. Distributive (four items) and procedural (seven items) justice were measured with Colquitt's (2001) organizational justice scale. Participants were instructed to rate the extent they experienced the described fair treatment on a scale ranging from 1 (to a very small extent) to 5 (to a very large extent). Sample items include "Are those outcomes justified given your performance?" ([[alpha].sub.1] = 0.95; [[alpha].sub.2] = 0.94) for distributive justice and "Have those procedures been applied consistently?" ([[alpha].sub.1]--0.91; [[alpha].sub.2] = 0.90) for procedural justice.

Perceived Organizational Support. Employees' support perceptions were assessed with the eight-item short-form of Eisenberger et al.'s (1986) survey of perceived organizational support. The survey captures the valuation of employees' contributions to the organization and their sense the organization looks out for their well-being (Rhoades and Eisenberger, 2002). A sample item is "The organization values my contribution to its well-being" ([[alpha].sub.1] = 0.92; [[alpha].sub.2] = 0.92).

Organizational Identification. Employees' organizational identity was assessed using Mael and Ashforth's (1992) five-item scale. The scale items indicate the extent to which employees agree with statements relating to their organizational psychological connection. A sample item is "When someone outside criticizes my company, it feels like a personal insult" ([[alpha].sub.1] = 0.82; [[alpha].sub.2] = 0.82).

Job Engagement. This motivational state was measured using nine of the 18 items from Rich et al.'s (2010) scale. Previous research used a shortened form of this scale (e.g., Haynie et al., 2016), providing some support for using a reduced number of items when assessing job engagement. Job engagement is meant to capture the physical, cognitive, and emotional energies invested into work and the nine items chosen were taken equally from these three components. A sample item is "I exert my full effort to my job" ([[alpha].sub.1] = 0.92; [[alpha].sub.2] = 0.93).

Control Variables. Job engagement is known to be influenced by supervisor characteristics and treatment (Christian et al., 2011). To account for these effects, interpersonal (four items: [[alpha].sub.1] = 0.95; [[alpha].sub.2] = 0.91) and informational (five items: [[alpha].sub.1] = 0.93; [[alpha].sub.2] = 0.89) justice, two interactional justice forms, were measured with Colquitt's (2001) organizational justice scale and used as controls. Because the purpose was to examine the impact of organizational forms of justice on job engagement, supervisors' fair treatment was controlled in order to extract the influence of more proximal leadership on employee motivation and to more fully test the influence of organizational forms of justice on job engagement.

RESULTS

Descriptive Statistics

Tables 1 and 2 present the descriptive statistics and correlations among the study variables from Samples 1 and 2, respectively. Variables related with each other in expected ways. For example, distributive and procedural justice positively correlated with POS and organizational identification while organizational identification positively correlated with job engagement. The correlations were examined pursuant to the suggestions by Bedeian (2014) and did not indicate any multicollinearity concerns. Although some correlations, such as the one between procedural justice and POS (r = 0.70 in Sample 1; r = 0.63 in Sample 2), were relatively strong, the strengths of these estimates do not exceed 0.70, excluding the correlation between interpersonal justice and informational justice (r = 0.66 in Sample 1; r = 0.73 in Sample 2), the two control variables. The high correlation between the controls was expected due to interpersonal and informational justice being two interactional justice forms (Colquitt, 2001), but not alarming because these variables are used as controls and not key constructs in the model. Additionally, the strengths of these associations are comparable to those found in meta-analytic results for organizational justice (Colquitt et al., 2013), POS (Rhoades and Eisenberger, 2002), and organizational identification (Riketta, 2005), and based on current standards (i.e., Bedeian, 2014) provide a sufficient indication of discriminate validity to proceed with examining the model's factor structure.

Confirmatory Factor Analysis

Prior to testing the hypotheses, the fit of the model structure was examined for both samples using Mplus software with estimations based on the default maximum likelihood parameter approach for continuous variables (Muthen and Muthen, 2007). Because the model contained a large number of parameter estimates approaching the minimally recommended 10:1 subject-to-item ratio (Bandalos, 2002) for structural equation modeling, scale parcels were used as factor indicators in the confirmatory factor analyses. Parceling combines scale items in order to reduce the number of factor indicators and improves the subject-to-item ratio. Tire parcels were created by sequentially averaging the highest and lowest loading items for each factor, which Landis et al. (2000) noted as a superior approach that best improves the reliability of the fit indices and communality of the factors (Little et al., 2013). Williams et al. (2009) in their guide to structural equation modeling argue scale parcels serve as more effective factor indicators when assessing latent constructs, so parceling appears to be a valid approach here. Using this approach, a specific number of parcels were created as factor indicators for each construct as follows: distributive justice (two parcels), procedural justice (three parcels), interpersonal justice (two parcels), informational justice (two parcels), POS (four parcels), organizational identification (two parcels), and job engagement (four parcels).

Using the parceled data for Sample 1 and allowing the exogenous variables to covary, the seven-factor model fit the data well: [chi square](131) = 215.98, p < 0.001, comparative fit index (CFI) = 0.98, Tucker-Lewis index (TLI) = 0.98, root mean square error of approximation (RMSEA) = 0.05, and standardized root mean squared residual (SRMR) = 0.04, with all factor loadings significant. Due to strong relationships between justice facets and the similarity of the mediator variables, two additional factor structures were analyzed. In the first model, distributive and procedural justice were combined as one factor, which produced poorer model fit, [chi square] (137) = 661.73, p < 0.001, CFI = 0.90, TLI = 0.87, RMSEA = 0.11, and SRMR = 0.05, and this model was significantly different than the seven-factor structure, [DELTA] [chi square] (6) = 445.75, p < 0.001. The second model combined POS and organizational identification into one factor, producing poorer model fit, [chi square] (137) = 376.39,p < 0.001, CFI = 0.95, TLI = 0.94, RMSEA = 0.08, and SRMR = 0.05, which was significantly worse than the seven-factor structure, [DELTA] [chi square] (6) = 160.41, p < 0.001.

In Sample 2, the seven-factor model fit the data well: [chi square] (131) = 266.81, p < 0.001, CFI = 0.98, TLI = 0.98, RMSEA = 0.05, and SRMR = 0.03, with all factor loadings significant. Again, distributive and procedural justice were combined into one factor, which produced poorer model fit, [chi square] (137) = 800.02,p < 0.001, CFI = 0.91, TLI = 0.89, RMSEA = 0.10, and SRMR = 0.04, and this structure was significantly worse than the seven-factor structure, [DELTA] [chi square] (6) = 533.21, p < 0.001. POS and organizational identification were also combined into one factor, producing poorer model fit, [chi square] (137) = 434.26, p < 0.001, CFI = 0.96, TLI = 0.95, RMSEA = 0.07, and SRMR = 0.05, which was significantly worse than the seven-factor structure, [DELTA] [chi square] (6) = 167.45, p < 0.001. Based on the fit of the hypothesized structure and the model comparisons across both samples, the hypotheses were tested using the proposed factor structure.

Hypotheses Testing

Hypothesis 1 proposed organizational identification mediates the relationships of (a) distributive and (b) procedural justice with job engagement. The model fit the data well from Sample 1 ([chi square] (77) = 108.30, p = 0.01, CFI = 0.99, TLI = 0.99, RMSEA = 0.04, and SRMR = 0.03) as well as Sample 2 ([chi square] (77) = 180.15, p < 0.001, CFI = 0.98, TLI = 0.98, RMSEA = 0.05, and SRMR = 0.04). The unstandardized coefficients are shown in Figure I.

The indirect effects of distributive and procedural justice with job engagement via organizational identification were then examined. Significant indirect effects are indicative of mediation (Sobel, 1982). Because indirect effects tend to violate normality, it is recommended to use a resampling method such as bootstrapping (MacKinnon et al., 2002). One thousand bootstrapped subsets from each sample were used to estimate 95% bias-corrected confidence intervals for the indirect effects. The indirect effects for distributive justice were significant in Sample 1 (Indirect effect = 0.06, [CI.sub.95%] [0.02, 0.12]) and Sample 2 (Indirect effect = 0.05, [CI.sub.95%] [0.01, 0.09]). Likewise, the indirect effects for procedural justice were significant in Sample 1 (Indirect effect = 0.09, [CI.sub.95%][0.04, 0.15]) and Sample 2 (Indirect effect = 0.14, [CI.sub.95%] [0.08, 0.23]). Hypothesis 1 was thus supported.

Perceived organizational support was introduced into the structural model to test Hypotheses 2 and 3. The model fit the data well for both Sample 1, ([chi square](135) = 222.19, p < 0.001, CFI = 0.98, TLI = 0.98, RMSEA = 0.05, and SRMR = 0.04), as well as Sample 2, ([chi square] (135) = 332.74, p < 0.001, CFI = 0.97, TLI = 0.97, RMSEA = 0.06, and SRMR = 0.06). As a way of judging the importance of POS in the process model, all paths with POS were constrained to zero to assess any potential detriment to model fit compared to the model including those paths. These constraints lead to poorer model fit in Sample 1, ([chi square] (139) = 521.50, p < 0.001, CFI = 0.93, TLI = 0.91, RMSEA = 0.10, and SRMR = 0.24) which was significantly different than the model including those paths, [DELTA] [chi square] (4) = 299.31, p < 0.001. Similar results were found in Sample 2 ([chi square] (139) = 697.01, p < 0.001, CFI = 0.93, TLI = 0.91, RMSEA = 0.09, and SRMR = 0.26) and this model was significantly different than the one with POS paths, [DELTA] [chi square] (4) = 474.82, p < 0.001. These results further support examining indirect effects associated with POS, so the authors proceeded with further indirect effect analyses. Figure II displays the unstandardized coefficients from both samples.

Bootstrapping was again used to estimate 95% bias-corrected confidence intervals to test the indirect effects for Hypotheses 2 and 3. Hypothesis 2 proposed POS would mediate the relationships of (a) distributive and (b) procedural justice with organizational identification. The indirect effect for distributive justice was significant in both Sample 1 (Indirect effect = 0.14, [CI.sub.95%] [0.08, 0.23]) and Sample 2 (Indirect effect = 0.11, [CI.sub.95%] [0.04, 0.19]). Additionally, the indirect effect for procedural justice was significant in Sample 1 (Indirect effect = 0.28, [CI.sub.95%] [0.18, 0.44]) and Sample 2 (Indirect effect = 0.33, [CI.sub.95%] [0.20, 0.50]), providing support for Hypothesis 2.

Hypothesis 3 argued the relationships of (a) distributive and (b) procedural justice with job engagement are indirectly related serially via POS and then organizational identification. The indirect effects for this dual-mediation were examined. The indirect effects for distributive justice in Sample 1 (Indirect effect = 0.03, [CI.sub.95%] [0.01, 0.07]) and Sample 2 (Indirect effect = 0.03, [CI.sub.95%] [0 01, 0.06]) were significant as well as for procedural justice in Sample 1 (Indirect effect = 0.07, [CI.sub.95%] [0.02, 0.14]) and Sample 2 (Indirect effect = 0.10, [CI.sub.95%] [0.05, 0.18]). Taken together, these results provide support for Hypothesis 3.

Because POS could still potentially directly impact job engagement while accounting for organizational identification effects, the justice-POS-job engagement indirect effect was also examined within the current model as a supplemental analysis. These indirect effects for distributive justice (Sample 1: Indirect effect = 0.06, [CI.sub.95%] [0.02, 0.16]; Sample 2: Indirect effect = 0.03, [CI.sub.95%] [0.00, 0.08]) and procedural justice (Sample 1: Indirect effect = 0.13, [CI.sub.95%] [0.04, 0.29]; Sample 2: Indirect effect = 0.08, [CI.sub.95%] [0.01, 0.18]) were significant across samples, indicating that organizational identification only partially mediates the relationship between POS and job engagement.

DISCUSSION

Organizational justice remains a popular topic of study due to its positive influence on job performance (Colquitt et al, 2001) and enhancing effects on motivational states such as job engagement (e.g., Haynie et al., 2016). Scholars recognize that motivation emerging from fair treatment may stem from desires to maintain favorable social exchange connections (Masterson et al., 2000) as well as social identities formed with the party providing the fair treatment (Tyler and Blader, 2003). Given these contrasting frameworks, a model that included both exchange-based (e.g., POS) and identity-based (e.g., OI) mechanisms through which justice impacts job engagement was tested. By examining the complementing role of these theories, the study responds to Van den Bos' (2005) call to test the relative contributions of justice's psychological explanations to advance knowledge of the simultaneous influence these theories have on employee reactions.

The current study extends upon prior findings by considering these mediating variables in unison and their likely ordering in the relationship between justice and job engagement. The results of the multi-sample study suggest that the effects of two key types of justice--distributive and procedural justice--on job engagement are enacted through a dual-mediated pathway. These justice perceptions heighten a social exchange relationship with the employing organization captured by POS which in turn increases identity formation as indicated by enhanced organizational identification and subsequent job engagement. The indirect influence of distributive and procedural justice with job engagement can thus be attributed to both social exchange and social identity mechanisms.

Limitations

As with any study, there are limitations to consider regarding the data collection process. First, the reliance on self-report scales potentially biases the relationships found among study variables (Podsakoff et al., 2012). These correlations, however, do not exceed thresholds of multicollinearity set by Brown (2014; i.e., 0.80) or Kline (2011; i.e., 0.85), and only the two control variables (interpersonal and informational justice) exceed the more conservative 0.70 threshold set by Bedeian (2014). Additionally, not all study variables are strongly correlated, suggesting that common method variance is unlikely (Brannick et al., 2010). This evidence, combined with the consistent findings across the homogenous depth of the participants in the first sample and the heterogeneous breadth of those from the second sample, suggest that common method variance is not of major concern.

Second, using a snowball data collection method, while offering benefits including the potential for a larger and more diverse sample than many traditional organizational field studies, has drawbacks, including non-randomness, increased likelihood for sample bias, and difficulty generalizing the findings (see Zinkhan et al., 1983). Additionally, collecting the measures at a single time period raises concerns over causality which is important when examining mediation models (Green et al., 2010). Given the cross-sectional nature of the data, Iacobucci et al.'s (2007) recommendation was adopted for assessing mediation or indirect effects in structural equation modeling, which encourages testing an alternate model in which causal order is reversed. Therefore, the hypothesized model was compared to one in which organizational identification predicted POS (justice[right arrow]organizational identification[right arrow]POS[right arrow]job engagement) instead of the originally proposed POS to organizational identification order. This alternative model's dual-mediated indirect effects were weaker in both Sample 1 for distributive (Indirect effect = 0.01, [CI.sub.95%][0.00, 0.04]) and procedural (Indirect effect = 0.02, [CI.sub.95%] [0.01, 0.06]) justice and in Sample 2 for distributive (Indirect effect = 0.01, [CI.sub.95%] [0.00, 0.02]) and procedural (Indirect effect = 0.02, [CI.sub.95%] [0.00, 0.06]) justice than the originally hypothesized order. Essentially, organizational identification appears to explain more of the reasons behind why POS promotes job engagement within justice-job engagement effects than does POS explaining why organizational identification promotes job engagement, lending some credence to the proposed causal order.

Theoretical and Practical Implications

By incorporating both a social exchange and social identity lens into justice-job engagement relations, support was found for the ordered mediating role of POS and organizational identification. Haynie et al. (2016) argued that both affect-based and relational mechanisms underlie the mediating effect of job engagement in justice-outcome relations. Through this study, additional understanding was highlighted regarding the reasoning employees use when deciding to become engaged from just treatment by examining the social exchange mechanisms underlying employees' organizational identity formation. Organizational identification has been shown to reflect the self-definitional aspects of organizational membership and social exchange processes presume a psychological distinction between the self and organization (Van Knippenberg and Sleebos, 2006). This research indicates, however, that it is through the behavioral clarity of greater social exchange as indicated by POS that organizational identification is created, leading to subsequent job engagement. There is strong support for the positive relationships of POS (Rhoades and Eisenberger, 2002) and organizational identification (Riketta, 2005) with many job performance behaviors likely requiring job engagement, so acknowledging the shared influence of these factors helps guide future research in these areas.

Although the indirect influence of justice-POS-organizational identification-job engagement was supported in both samples, the indirect effect of justice-POS-job engagement still remained significant. POS has been argued to capture the social exchange relationship formed with the employing organization (e.g., Masterson et al., 2000) but it also is thought to represent value expressive information meeting employees' socioemotional needs (Rhoades and Eisenberger, 2002). Future research is thus needed to examine what degree social exchange norms, socioemotional needs, or both explain the justice-POS-organizational identification-job engagement relationship and how these factors may provide further understanding of the reasons POS still positively enhances job engagement within this indirect process model.

Lastly, the findings suggest the importance of considering social exchange in the group engagement model. Of note, the results occurred while controlling for the impact of interpersonal justice and informational justice on job engagement, thus demonstrating the extent distributive and procedural justice truly relate with job engagement via POS and organizational identification. According to the group engagement model, distributive and procedural justice offer the behavioral guidance and consistent treatment necessary for meeting work goals so employees are willing to generate group identities and pursue desired outcomes (Tyler and Blader, 2003). These expected behaviors likely captured in job engagement emerge from social identities to the extent norms prescribe such behaviors (Van Knippenberg and Ellemers, 2003). These normative expectations are thus captured by strong social exchange relationships which serves as rationale for the causal ordering, which aligns with research suggesting this directional effect (Edwards and Peccei, 2007). Thus, it would benefit future group engagement and more broadly identity-related research by acknowledging the influence of social exchange tenets in their theoretical models.

For practitioners, the results further support the importance of organizational justice for promoting job engagement among employees. By offering equitable pay and establishing fair policies and procedures that are consistent, free of bias, and ethical (Colquitt et al., 2005), organizational leaders can promote job engagement. Microsociological researchers studying identities have long recognized the external influences on an individual's identity such that the expectations others hold help shape and reinforce individual identities (Burke, 1991). As such, identities are not static so external forces can both enhance (Burke, 1991) and deter (Farmer et al, 2003) their creation. Organizational leaders thus need to be aware of the fairness guiding organizational rules and policies in order to more fully gain the motivation of their employees (Ouchi, 1980).

CONCLUSION

The examination of the extent justice-job engagement effects are explained by a dual-sequence of mediators such that a social exchange relationship precedes social identity within a motivational process model. These findings increase the understanding from prior researchers that have looked at the two theoretical perspectives as independent mediators of the justice-job engagement effect. Future research should consider both social exchange- and identity-mechanisms when examining the motivation underlying justice effects.

Jeffrey J. Haynie

Assistant Professor of Management

Louisiana Tech University

jhayni1@gmail.com

C. Brian Flynn

Assistant Professor of Management

University of North Florida

brian.flynn@unf.edu

John E. Baur

Assistant Professor of Management

University of Nevada, Las Vegas

john.baur@unlv.edu

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(1) Correspondence about this manuscript should be sent to Jeffrey J. Haynie, P. O. Box 10318, Department of Management, Ruston, LA 71272. Email may be sent tojhaynil@gmail.com.

Caption: Figure I Unstandardized path coefficients from hypothesized model excluding perceived organizational support

Caption: Figure II Un standardized path coefficients from hypothesized model including perceived organizational support
Table 1
Means, Standard Deviations, Reliabilities, and
Correlations for Variables in Sample 1

                                      Mean   S.D.      1

1. Distributive Justice               3.67   0.90   (0.95)
2. Procedural Justice                 3.42   0.77   0.62 ***
3. Interpersonal Justice              4.53   0.68   0.40 ***
4. Informational Justice              3.93   0.89   0.43 ***
5. Perceived Organizational Support   3.92   0.69   0.63 ***
6. Organizational Identification      3.75   0.65   0.41 ***
7. Job Engagement                     4.38   0.51   0.27 ***

                                         2          3          4

1. Distributive Justice
2. Procedural Justice                 (0.91)
3. Interpersonal Justice              0.42 ***   (0.95)
4. Informational Justice              0.50 ***   0.66 ***   (0.93)
5. Perceived Organizational Support   0.70 ***   0.35 ***   0.47 ***
6. Organizational Identification      0.40 ***   0.26 ***   0.29 ***
7. Job Engagement                     0.32 ***   0.25 ***   0.21 ***

                                         5          6         7

1. Distributive Justice
2. Procedural Justice
3. Interpersonal Justice
4. Informational Justice
5. Perceived Organizational Support   (0.92)
6. Organizational Identification      0.55 ***   (0.82)
7. Job Engagement                     0.40 ***   0.50 ***   (0.92)

Note: Sample 1 n = 308; Cronbach's alphas
are found on the diagonal in parentheses.

* p < 0.05; ** p <0.01; *** p < 0.001.

Table 2
Means, Standard Deviations, Reliabilities, and
Correlations for Variables in Sample 2

                                      Mean   S.D.   1

1. Distributive Justice               3.65   0.97   (0.94)
2. Procedural Justice                 3.41   0.83   0.60 ***
3. Interpersonal Justice              4.28   0.82   0.40 ***
4. Informational Justice              3.91   0.86   0.45 ***
5. Perceived Organizational Support   3.71   0.84   0.57 ***
6. Organizational Identification      3.39   0.84   0.37 ***
7. Job Engagement                     4.04   0.68   0.33 ***

                                      2          3          4

1. Distributive Justice
2. Procedural Justice                 (0.90)
3. Interpersonal Justice              0.45 ***   (0.91)
4. Informational Justice              0.50 ***   0.73 ***   (0.89)
5. Perceived Organizational Support   0.63 ***   0.55 ***   0.60 ***
6. Organizational Identification      0.41 ***   0.37 ***   0.38 ***
7. Job Engagement                     0.44 ***   0.37 ***   0.39 ***

                                      5          6          7

1. Distributive Justice
2. Procedural Justice
3. Interpersonal Justice
4. Informational Justice
5. Perceived Organizational Support   (0.92)
6. Organizational Identification      0.55 ***   (0.82)
7. Job Engagement                     0.52 ***   0.52 ***   (0.93)

Note: Sample 2 n = 460; Cronbach's alphas
are found on the diagonal in parentheses.

* p < 0.05; ** p < 0.01; *** p < 0.001.
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Author:Haynie, Jeffrey J.; Flynn, C. Brian; Baur, John E.
Publication:Journal of Managerial Issues
Article Type:Report
Date:Mar 22, 2019
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