A TECHNIQUE TO MEASURE COLLEGE STUDENTS ON THE DEPRESSION- ELATION CONTINUUM.
Depression can be considered a very serious disorder that affects not only the victims, but also the friends and family involved. About 16% of American adults will be clinically depressed at some points in their life.
Over the past few decades there has been a movement towards improving the measurement of depression so that research, clinical practice, and the overall understanding of this construct could be enhanced. This impetus has sparked the need for psychometric instruments that allow for a quick assessment of depressive symptoms, with the self-report method hitherto the most efficient way to quantify depression (Bech, 1992; Beckham & Leber, 1995; Gotlib & Cane, 1989; Gotlib & Hammen, 1992; Kate, Shaw, Vallis, & Kaiser, 1995). While more than 280 self-report scales have been constructed for measuring depression since 1918, the Beck Depression Inventory (BDI; Beck, Rush, Shaw, & Emery, 1979; Beck, Ward, Mendelson, Mock, & Erbaugh, 1961), the Center for Epidemiological Studies Depression Scale (CES-D; Radloff, 1977), the Hamilton Rating Scale for Depression (HRSD; Hamilton, 1960), and the Zung Self-Rating Depression Scale (SDS; Zung, 1965) have been four of the most widely used depression measurement tools (Santor, Gregus, & Welch, 2006; Shafer, 2006).
The majority of studies on depression have come from investigating analogue (e.g., university students) samples using self-report inventories (Tennen, Hall, & Affleck, 1995; Vredenburg, Flett, & Krames, 1993). Of the many self-report inventories available, different editions of the BDI appear to be the most popular scales used in nonclinical samples (Ruscio & Ruscio, 2002). The BDI and its revisions (Beck et al., 1996; Beck et al., 1979; Beck et al., 1961) have been heralded as some of the best self-report measures of depression (Hiroe et al.,2005; Richter et al., 1998; Shafer, 2006), used extensively (Whisman et al., 2013; Beck, Steer, & Garbub, 1988; Katz et al., 1995; Scogin et al., 1988), and translated into many languages for cross-cultural studies (Dere et al., 2014, Richter et al., 1998; Wu, & Huang, 2014).
Beck constructed the original BDI's items by observing symptoms that were common among depressed patients and uncommon among non-depressed individuals. The categories of symptoms that the BDI sought to assess were the individuals' perceived emotion status, their awareness of behavioural actions, and their perceived somatic symptoms. The most recent edition of the Beck Depression Inventory (BDI-II; Beck et al., 1996) was formed so that the BDI's items were commensurate with the current criteria for major depressive disorder, as made explicit in the Diagnostic and Statistical Manual of Mental Disorders-IV (Association, 1994).
Beck et al. (1996) have claimed that the BDI does not represent any particular theory of depression, and that it simply reflects the observed symptoms of persons who are depressed. This explains why the BDI's originally developed usage was to assess the severity of depression once a diagnosis of depression had already been made (Beck et al., 1961; Williams, 1992). Despite this originally intended usage, the scale has frequently been used for the detection and quantification of depression in nonclinical populations (Hatzenbuehler, Parpal, & Matthews, 1983). Several researchers (e.g., Gotlib, 1984; Hatzenbuehler et al., 1983; Lips & Ng, 1985; Santor, Ramsay, & Zuroff, 1994) have questioned whether the application of the clinically inspired BDI to nonclinical populations is justifiable, as there are many issues that arise in using a scale that was created for a specific population (i.e., clinically diagnosed depressed individuals) on a different population, such as a decrease in the reliability and validity of the measuring instrument.
Recent examinations of the BDI-II have suggested the decrease of validity may be due to its incapability to discriminate between disorders (Krueger & Markon, 2006; Subica et al., 2014) attributed to the significant di agnostic overlap (Krueger & Markon, 2006) amongst internalizing disorders (e.g. anxiety, depression, phobia) (Subica et al., 2014).
Other researchers have questioned the suitability of using instruments that measure uni-polar constructs (such as depression) in nonclinical populations. In their article relating to how practicing researchers should construct and evaluate psychological tests, Reise, Waller, and Comrey (2000) have suggested that the measurement of a construct such as uni-polar depression may not be appropriate for non clinical and undergraduate student samples, since these scales are not well constructed to measure individual differences effectively. The BDI instruments are prime examples of this objection, as can be seen by way of a statistical 'flooring' effect so common among nonclinical studies. Although there is potential for each item to range from 0 to 3, studies in nonclinical (e.g., student) populations typically report av erage scores below 1, and even for psychiatric samples the mean item scores rarely exceed 2 (Richter et al., 1998; Santor et al., 1994). This in turn produces a strong positive skewness when scores are aggregated, and requires the data to be transformed before many parametric and nonparamerric statistical techniques could be performed (Tabachnick & Fidell, 2013; Whisman et al., 2013).
Another concern regarding the use of the BDI on nonclinical samples is the general interpretation of what may be considered a response of absent or neutral symptomology. Specifically, nonclinical populations may re quire response options that more accurately reflect their lived experiences, both positive and negative, rather than having to settle for the neutral option available in the BDI. Joseph et al. (2006) have questioned whether the BDI should include additional response options to create a more integrative body of knowledge in regards to the study of depression and its elation counterpart.
Positive psychology and bipolarity
In recent years there has been a movement in the field of positive psychology towards creating psychometric instruments that mea sure the full spectrum of various constructs (i.e., happiness and sadness) for research purposes and for clinicians to utilize in therapeutic settings (e.g., Joseph & Linley, 2004; Keyes & Lopez, 2002; Maddux, 2002). Positive psychologists are concerned with the pro motion of general well-being as well as the alleviation of more specific psychopathology, and there has been growing interest in this framework among practitioners (e.g., Linley & Joseph, 2004; Lopez & Snyder, 2003; Snyder & Lopez, 2002). Researchers have noted that an instrument that is able to simultane ously measure elation and depression may potentially be useful in both therapeutic and research settings (see, for example, Maddux, Snyder, & Lopez, 2004; Ruini & Fava, 2004; Seligman, 2002). With respect to research, instruments that encompass a more complete range of human emotions may be more reliable and valid than scales assessing a truncated construct, as the participants will not feel forced to choose an option that may not be representative of their current state. In clinical practice, it may be beneficial for clinicians to have an instrument that is more able to assess related changes across a continuum.
With the intent of amending the BDI-II to improve its reliability and validity when administering the scale to nonclinical pop ulations, a survey package consisting of 19 positive items that semantically reflected the response options in the BDI-II (1) was created.
(1) Permission to administer a modified version of the BDI-II was obtained from Pearson Inc.. Reference Number: 6596089-0-963903.
These added response options mirrored the three negative scenario options in the original BDI-2 as shown in Table 1. Both items 16: changes in sleeping pattern and 18: change in appetite were not included in the package.
The current study examines how under-graduate respondents--representative of a nonclinical population--would endorse the reflected response options in a separate 19-item scale. It was thought that this would allow information from items assessing elation to be gathered alongside the BDI-1I, but with out directly affecting the BDI-II's original metric. This approach would also avoid much of the theoretical ambiguities that may be inherited when mixing positive and negative item content.
A sample of 343 undergraduate students were asked to voluntarily fill out a survey package. This sample was composed of 246 females and 95 males, where two participants did not disclose their gender. The overall mean and median age for the 343 participants were 21.52 (S D = 8.05) and 19, respectively, where the males ([M.sub.age] = 23.69, median = 20) were significantly older than the females ([M.sub.age] = 20.85, median = 19; t(339) = 3.01, p < .003).
Participants were asked to complete a survey package consisting of the 21-item BDI-II scale and an attached scale that consisted of the 19 positive items created with semantical ly reflected response options. The composite scores of this package ranges from 0 to -57. Values with greater absolute magnitude are indicative of higher levels of elation. Written and verbal instructions were given to participants not to look back at their previously answered form to determine the responses that had already been given. Half of the packages began with the BDI-II first followed by the reflected scale, and the other half ended with the BDI-II. No ordering effect was detected.
Design & Procedures
Respondents completed the survey pack age voluntarily, and no monetary incentive was provided to complete the questionnaires. Guttman-Cronbach [alpha]'s (Cronbach, 1951; Guttman, 1945) were computed for both the BDI-II and the reflected scale within the administered packages to estimate internal consistency. Exploratory principle axis analyses were conducted using polychoric correlations to examine the general factor structure of the two scales, and scree plots were examined to estimate the number of potential factors us ing Cattell's (1966) and the parallel analysis (Horn, 1965) criteria. Parallel analysis is now often recommended as the best method to as sess the true number of factors (Lance, Butts, & Michels, 2006; Velicer, Eaton, & Fava, 2000), and can help augment the interpreta tion of a scree plot (Drasgow & Lissak, 1983; Horn, 1965; Longman, Cota, Holden, & Fek ken, 1989; Montanelli & Humphreys, 1976; Zwick & Velicer, 1986). The BDI-II and the 19-item scale reflecting the polar opposite response options were then combined to create a statistically bipolar scale. Once combined, the same statistical analyses were performed to determine the reliability and validity of the combined scale.
Results Individual Instruments
The BDI-II ([alpha] = .916, 95% CI [.888, .935]) and the reflected scale ([alpha] = .888, 95% CI [.865, .906]) possessed high estimates of Guttman-Cronbach alphas, indicative of statistical homogeneity. To more directly compare the two instruments' reliability estimates the Spearman-Brown prophecy formula was utilized to estimate how the 19-item scale [alpha] might have been increased had there been 21 items. The formula increased the reliability estimate from [alpha] = .888 to [[alpha].sup.0] = .897.
The constructed scree plots revealed that both scales were dominated by a single factor, with potentially an additional smaller factor that might have been useful for extraction. The Parallel analyses confirmed the observation that two factors should be extracted for both instruments. The BDI-II's extracted eigenvalues accounted for 50.14% and 4.72% of the scales variance, while the reflected scale's eigenvalues initially accounted for 41.92% and 5.03% of the variance, respectively. A strong initial factor was also made evident by the positive and fairly high corrected item-total (CIT) correlations in both scales (see Table 2), with the BDI-II (M = .57, SD = .10) possessing a slightly higher mean CIT correlation than the reflected scale (M = .50, SD = .07). Given the relatively high Guttman-Cronbach as and CIT correlations, as well as evidence for a single dominant eigenvalue in both instruments, it was deemed provisionally acceptable to aggregate items in each instrument to create composite scores. The BDI-II's composite score had a mean of 9.36 (SD = 8.49) and, as expected, displayed a strong positive skewness (see Figure 1; Skew ness = 1.751, SE = 0.132). The mean score for the reflected scale was -17.95 (SD = 11.56), and as seen in Figure 1 possessed a moderate negative skewness (Skewness = -0.634, SE = 0.132). Additionally, the zero-order correlation coefficient between these two composite scores was found to be -.583 ( p < .001 ) (2).
The Statistically Bipolar Instrument
The next approach in the analysis was to combine these two scales and treat them as a single instrument. This was done to examine the effects these new 19-items would have with respect to the original BDI-II items and to observe how the psychometric properties would change for this statistically bipolar scale. Guttman-Cronbach's [alpha] was again found to be quite high ([alpha] = .925, 95% CI [.909, .937]), and the CIT correlations were all positive and--aside for the reflected item 21--above .3 (M = .48, SD = .09; see Table 2). A composite score was then computed and can be seen in Figure 2. This composite score appeared to be approximately normally distributed about a mean score of-8.48 (SD = 17.38; Skewness = 0.288, SE = 0.132; Kurtosis = 0.389, SE = 0.263).
A scree plot was then built to estimate the number of underlying factors using the same criteria as before. Visually it appeared that two or three factors would be sufficient for extraction, and the accompanying parallel analysis suggested that three factors should be extracted. However, after extraction and rotation there did not appear to be any loadings above .33 for the third factor, therefore a two-factor structure was used instead. These two factors were extracted and rotated using the same criteria as before, and the results are displayed in Table 2. The extraction results revealed comparable rotated sums of squares (9.75 and 8.97, respectively), and the pattern matrix identified every item with their respective instrument. Factor 1 possessed all the BD1-II items with 1 cross-loading (.38; item 2) while factor 2 possessed all the reflected items and also had one cross-loading (.35; item 21). These oblique factors were related by a coefficient of .65.
Individually the two scales demonstrated high scale score reliability and evidence for a dominant factor, and when considered to gether, the full-spectrum scale appeared to be measuring a common higher order construct by evidence of the relatively high factor correlations and the positive CIT correlations. The average composite score for the BDI-II was within the range expected for a non-clinical sample, and demonstrated its characteristically strong positive skewness. The average composite score for the semantically reflected scale was well below the 'absence of elation' score of 0, and when both scales were considered together the composite score revealed that there was a higher endorsement of positive self-reported symptoms than nega tive. This despite the fact that there were two more items relating to depression rather than elation (21 versus 19), and hence had a stronger impact on the scales' composite score. Additionally, the statistically bipolar scale transformed the previously skewed distributions into an apparently normal distribution.
The results displayed an apparent improvement to the BDI-II's classification system. The BDI-II classified 20.4% (n = 129) of the participants as possessing either a mild, moderate or severe degree of depression, with the majority of these classifications (n = 67) appearing in the 'mildly depressed' range. However, using the same classification scheme the bipolar scale only classified 6.8% (n = 47) of the participants as possessing a mild to severe degree of depression. This large drop makes conceptual sense considering the population sampled. However, we must keep in mind an obvious limitation of this study in that females were predominantly sampled!
Depression and elation are so strongly related that it is not justifiable to leave either complementary pole out of measurement instruments--especially if they are ever to be used in both clinical and nonclinical settings. Many psychological tests seeking to measure only one pole of an ostensibly bipolar construct would likely benefit by being paired with their complementary counterpart, yielding more meaningful results with regards to the nature of these uniquely bipolar continuums. As this study has demonstrated, the meaningfulness of the BDI-II's total score could be enhanced when compared alongside a comparable 'elation' based scale, which is more appropriate for more holistically understanding nonclinical as well as clinical populations. This is hopefully the first of many steps to expand the under standing of these uniquely bipolar constructs, and shows promise for moving psychological research away from using specialized hospital scales to using instruments appropriate for clinical as well as nonclinical individuals.
This study was supported by the Nipissing University Research Grant, Reference Number: 1523:7100.
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Department of Psychology, Nipissing University
R. PHILIP CHALMERS
Department of Educational Psychology, University of Georgia
DEBORAH M. FLYNN
Department of Psychology, Nipissing University
ADAM J. McLANDRESS
Department of Psychology, Nipissing University
VICTORIA G. L. STEADMAN
Department of Psychology, Nipissing University
(2) The correlation between these two composites was expected to be positive since the scoring of the reflected scale is based on a descending metric, beginning at zero. Therefore values with greater magnitude on the reflected scale should be associated with lower values on the BDI-II.
Table 1. Using BDI-ll's item 1 as an example to show the associated semantically reflected options as well as how they are scored. BDI-II 1. Happiness/Sadness Scoring to keep the BDI-II's original metric 0 I do not feel sad. 1 I feel sad much of the time. 2 I am sad all the time. 3 I am so sad or unhappy that I can't stand it. BDI-II The Reflected Positive Scale Scoring Scoring opposite to the BDI-II's metric 0 I do not feel sad. 1 I am happy much of the time. 2 I am happy all the time. 3 I could not possibly be happier. Table 2. Corrected-item-total (CIT) correlations and a principle axis factor extraction with a Promax (k = 4) rotation when the two instruments were treated as a whole. Since there were minimal cross-factor loadings, the 19 reflected items could be compared to their semantically opposite counterparts located within the BDI-II (excluding items 16 and 18 since they were not reflected). Factor Pattern Loadings Items BDI-II 1 (Sadness/Hapiness) .55 2 (Pessimism/Optimism) .41 3 (Failure/Success) .60 4 (Lost/Gained Pleasure) .61 5 (Guilty/Guilt-free) .72 6 (Punishment/Reward) .98 7 (Self-Dislike/-Like) .76 8 (More/Less Self-Critical) .77 9 (Suicidal/Pro-life) .57 10 (Uncontrolled/Controlled Crying) .35 11 (Agitation/Relaxation) .64 12 (Lost/Gained Interest) .51 13 (Indecisive/Decisive) .63 14 (Worthless/Worthy) .93 15 (Lost/Gained Energy) .48 16 (Sleep Changes) .38 17 (More/Less Irritable) .57 18 (Appetite Changes) .70 19 (Worse/Better Concentration) .84 20 (Tired/Alert) .68 21 (Less/More Libido) .33 Eigenvalues 15.89 Rotated SS 9.75 Items Reflected Scale 1 (Sadness/Hapiness) .58 2 (Pessimism/Optimism) .60 3 (Failure/Success) .55 4 (Lost/Gained Pleasure) .73 5 (Guilty/Guilt-free) .30 6 (Punishment/Reward) .54 7 (Self-Dislike/-Like) .53 8 (More/Less Self-Critical) .67 9 (Suicidal/Pro-life) .46 10 (Uncontrolled/Controlled Crying) .47 11 (Agitation/Relaxation) .76 12 (Lost/Gained Interest) .72 13 (Indecisive/Decisive) .72 14 (Worthless/Worthy) .57 15 (Lost/Gained Energy) .76 16 (Sleep Changes) - 17 (More/Less Irritable) .67 18 (Appetite Changes) - 19 (Worse/Better Concentration) .66 20 (Tired/Alert) .74 21 (Less/More Libido) .62 Eigenvalues 2.83 Rotated SS 8.97 CIT Correlations Items BDI-II 1 (Sadness/Hapiness) .47 2 (Pessimism/Optimism) .54 3 (Failure/Success) .44 4 (Lost/Gained Pleasure) .57 5 (Guilty/Guilt-free) .43 6 (Punishment/Reward) .34 7 (Self-Dislike/-Like) .58 8 (More/Less Self-Critical) .42 9 (Suicidal/Pro-life) .50 10 (Uncontrolled/Controlled Crying) .42 11 (Agitation/Relaxation) .54 12 (Lost/Gained Interest) .57 13 (Indecisive/Decisive) .44 14 (Worthless/Worthy) .47 15 (Lost/Gained Energy) .53 16 (Sleep Changes) .38 17 (More/Less Irritable) .57 18 (Appetite Changes) .36 19 (Worse/Better Concentration) .55 20 (Tired/Alert) .57 21 (Less/More Libido) .31 Eigenvalues - Rotated SS - Items Reflected Scale 1 (Sadness/Hapiness) .56 2 (Pessimism/Optimism) .50 3 (Failure/Success) .50 4 (Lost/Gained Pleasure) .35 5 (Guilty/Guilt-free) .45 6 (Punishment/Reward) .54 7 (Self-Dislike/-Like) .54 8 (More/Less Self-Critical) .47 9 (Suicidal/Pro-life) .58 10 (Uncontrolled/Controlled Crying) .43 11 (Agitation/Relaxation) .49 12 (Lost/Gained Interest) .57 13 (Indecisive/Decisive) .44 14 (Worthless/Worthy) .54 15 (Lost/Gained Energy) .47 16 (Sleep Changes) - 17 (More/Less Irritable) .52 18 (Appetite Changes) - 19 (Worse/Better Concentration) .54 20 (Tired/Alert) .51 21 (Less/More Libido) .17 Eigenvalues - Rotated SS -
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|Author:||Chow, Peter; Chalmers, R. Philip; Flynn, Deborah M.; McLandress, Adam J.; Steadman, Victoria G.L.|
|Publication:||College Student Journal|
|Article Type:||Brief article|
|Date:||Jun 22, 2018|
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