Center for Epidemiologic Studies Depression Scale Review and Revision (Cesd and Cesd-r)
PLoS One. 2013; 8(3): e58067.
The Center for Epidemiologic Studies Depression Scale: A Review with a Theoretical and Empirical Examination of Item Content and Gene Structure
R. Nicholas Carleton
1 The Anxiety and Disease Behaviour Laboratory, Department of Psychology, Academy of Regina, Regina, Saskatchewan, Canada,
Michel A. Thibodeau
1 The Anxiety and Illness Behaviour Laboratory, Department of Psychology, Academy of Regina, Regina, Saskatchewan, Canada,
Michelle J. North. Teale
1 The Anxiety and Illness Behaviour Laboratory, Section of Psychology, University of Regina, Regina, Saskatchewan, Canada,
Patrick G. Welch
i The Anxiety and Illness Behaviour Laboratory, Section of Psychology, Academy of Regina, Regina, Saskatchewan, Canada,
Murray P. Abrams
i The Anxiety and Affliction Behaviour Laboratory, Department of Psychology, University of Regina, Regina, Saskatchewan, Canada,
Thomas Robinson
2 Regina Qu'Appelle Health Region, Functional Rehabilitation Program, Regina, Saskatchewan, Canada,
Gordon J. Thousand. Asmundson
ane The Feet and Illness Behaviour Laboratory, Section of Psychology, Academy of Regina, Regina, Saskatchewan, Canada,
Hamid Reza Baradaran, Editor
Received 2012 Oct 17; Accepted 2013 January 29.
Abstract
Background
The Middle for Epidemiologic Studies Depression Scale (CES-D; Radloff, 1977) is a ordinarily used freely available self-written report measure of depressive symptoms. Despite its popularity, several contempo investigations have called into question the robustness and suitability of the commonly used 4-cistron 20-item CES-D model. The goal of the current study was to address these concerns by confirming the factorial validity of the CES-D.
Methods and Findings
Differential particular functioning estimates were used to examine sexual practice biases in item responses, and confirmatory factor analyses were used to assess prior CES-D factor structures and new models heeding current theoretical and empirical considerations. Data used for the analyses included undergraduate (n = 948; 74% women), community (due north = 254; 71% women), rehabilitation (due north = 522; 53% women), clinical (n = 84; 77% women), and National Health and Nutrition Test Survey (NHANES; n = 2814; 56% women) samples. Differential item functioning identified an detail as inflating CES-D scores in women. Comprehensive comparison of the several models supported a novel, psychometrically robust, and unbiased 3-cistron fourteen-item solution, with factors (i.e., negative bear upon, anhedonia, and somatic symptoms) that are more in line with electric current diagnostic criteria for depression.
Conclusions
Researchers and practitioners may benefit from using the novel gene structure of the CES-D and from beingness cautious in interpreting results from the originally proposed scale. Comprehensive results, implications, and future enquiry directions are discussed.
Introduction
The Diagnostic and Statistical Manual of Mental Disorders, Quaternary Edition, Text Revision [1] characterizes low as a multidimensional construct comprising negative emotion (i.eastward., negative affect; Criterion A1), an absence of positive emotions (i.eastward., anhedonia; Criterion A2), and a cluster of physical symptoms (i.due east., somatisation; Criteria A3-5). The Heart for Epidemiologic Studies Depression Scale (CES-D) [ii] is among the most popular measures of depressive symptoms, likely owing its popularity to being gratis and by and large comparable [three]–[5] with the well-established Beck Depression Inventories [6], [7]. Despite its popularity, the CES-D has areas of business organization, especially in its latent factor structure and item content.
The CES-D was originally posited as having a iv-gene structure representing depressed touch on, absence of positive touch or anhedonia, somatic activity or inactivity, and interpersonal challenges [2]. The CES-D items and structure were not designed a priori to reflect diagnostic criteria at the time of its evolution [8] and recent investigations take called into question the robustness and stability of the original iv-factor 20-item structure [9]–[xi]. Indeed, over 20 alternative factor solutions have been reported (Tabular array 1) and take suggested the presence of one, two, iii, and four factors [12]–[14]. The bulk of factor-analytic studies of the CES-D have employed principal component analysis with orthogonal rotation [4], an analytic approach that may accept theoretically improbable assumptions and biased factor solutions [15]. The shift away from such approaches is not a shift abroad from exploratory gene analyses, but a shift towards the best practices for such analyses; that said, exploratory factor analyses tend to be exploratory. In the case of constructs that are established (east.g., low), confirmatory factor analyses may be more informative as measures are designed to fit a construct, instead of naming constructs to fit the results from a measure.
Table ane
Prior multi-factorial model structures sorted by publication appointment.
| Factors (Items) | CES-D Item Number and Posited Gene Loading | ||||||||||||||||||||
| Reference | 1 | 2 | 3 | 4 | 5 | 6 | 7 | 8 | nine | x | 11 | 12 | 13 | 14 | xv | 16 | 17 | 18 | 19 | 20 | |
| Radloff, 1977 Model A [2] | 4 (xvi) | 1 | one | 2 | iv | 2 | 1 | 4 | one | 4 | 2 | 3 | iv | two | 2 | three | 1 | ||||
| Radloff, 1977 Model B [2] | four (20) | i | one | 2 | iv | 1 | 2 | 1 | 4 | 3 | 3 | 1 | 4 | 1 | 2 | 3 | 4 | 2 | 2 | 3 | 1 |
| Radloff, 1977 Model C [2]; Shafer, 2006 [4]; Williams, 2007 [xiv] | 4 (20) | one | 1 | 2 | 4 | ane | two | 1 | 4 | 2 | two | 1 | 4 | i | ii | 3 | 4 | 2 | ii | 3 | 1 |
| Burnam, 1988 [66]; Tuunainen, 2001[67] | 1 (half-dozen) | ane | i | 1 | one | i | 1 | ||||||||||||||
| Shrout, 1989 Model A [58] | ane (5) | 1 | 1 | i | one | 1 | |||||||||||||||
| Shrout, 1989 Model B [58] | i (five) | i | ane | 1 | ane | 1 | |||||||||||||||
| Radloff, 1991 [47] | 4 (17) | 1 | ane | two | iv | one | 2 | one | 4 | 1 | 4 | 2 | 3 | four | 2 | 2 | 3 | 1 | |||
| Kohout, 1993 [21]; Carpenter, 1998 [59]; Irwin, 1999 Model A [60] | 4 (19) | 1 | i | ii | 4 | 1 | 2 | 1 | 4 | iv | 1 | 4 | 1 | two | 3 | four | ii | 2 | 3 | 1 | |
| Kohout, 1993 [21]; Carpenter, 1998 [59]; Irwin, 1999 Model B [60] | four (eleven) | one | 2 | 1 | 1 | four | ii | 3 | 4 | 2 | iii | 1 | |||||||||
| Kohout, 1993 [21]; Carpenter,1998 [59]; Irwin, 1999 Model C [60] | iv (10) | i | 2 | ii | iv | one | 3 | 4 | 1 | three | 2 | ||||||||||
| Andresen, 1994 [40]; Cheng, 2006 [xvi] | 3 (10) | ane | 2 | i | ii | 3 | ane | ii | 3 | 1 | 2 | ||||||||||
| Santor, 1997[61]; Herrero, 2006 [62] | one (nine) | 1 | 1 | one | 1 | one | 1 | ane | 1 | i | |||||||||||
| Boey, 1999 Model A [63] | one (ten) | one | i | 1 | 1 | ane | one | one | 1 | 1 | 1 | ||||||||||
| Boey, 1999 Model B [63] | 2 (x) | one | 1 | 1 | 1 | 2 | ane | 1 | 2 | i | 1 | ||||||||||
| Rouch-Leroyer, 2000 [64] | 1 (5) | i | 1 | 1 | 1 | 1 | |||||||||||||||
| Schroevers, 2000 [10]; Rivera-Medina, 2010 [29] | 2 (20) | 1 | ane | 1 | ii | 1 | 1 | 1 | two | 1 | i | 1 | ii | 1 | 1 | i | 2 | 1 | 1 | one | 1 |
| Yen, 2000 [65] | 3 (17) | 1 | 1 | 4 | 1 | 1 | 1 | iv | 1 | 4 | 2 | 2 | 2 | iv | 2 | two | 2 | 1 | |||
| Bush-league, 2004 Both Sexes [68] | 4 (xx) | 1 | three | 1 | 4 | i | 1 | 1 | iv | 2 | one | 1 | four | iii | ii | two | 4 | 3 | ii | ii | 1 |
| Bush, 2004 Men But [68] | 3 (20) | ane | 1 | 2 | iii | 1 | 1 | 1 | 3 | 2 | 1 | ane | 3 | 2 | 2 | 2 | 3 | ii | 2 | 2 | 1 |
| Bush, 2004 Women Just [68] | 4 (20) | 1 | 3 | 1 | 4 | 1 | 1 | 3 | 4 | 1 | ane | 3 | iv | 3 | ii | 2 | 4 | 1 | 2 | 2 | 3 |
| Cole, 2004 [69] | four (10) | 1 | ii | 3 | 1 | i | iii | 4 | iv | 2 | iv | ||||||||||
| Stansbury, 2006 [11] | 3 (16) | 1 | one | ii | one | two | one | 3 | 2 | one | i | 2 | iii | 2 | 2 | three | 1 | ||||
| Lee AE, 2008 [22] | ii (10) | i | 1 | 1 | 1 | ii | i | 1 | 2 | i | i | ||||||||||
| Lee SW, 2008 Model A [thirteen] | 2 (16) | ane | one | 1 | 1 | 1 | 1 | 1 | 2 | 1 | 2 | 2 | 2 | 2 | two | 2 | 1 | ||||
| Lee SW, 2008 Model B [13] | 3 (16) | i | 1 | 2 | ane | ii | one | 3 | 3 | 1 | 1 | 2 | three | 2 | two | iii | 1 | ||||
Many researchers take also questioned the validity and psychometric properties of several items on the CES-D [16]–[25]. Items potentially assessing somatic concerns (e.g., "I felt that everything I did was an effort") may artificially inflate CES-D scores for elderly or chronic pain populations [26], [27]. 2 socially-focused items (i.due east., "People were unfriendly" and "I felt that people disliked me") are believed to potentially derange the validity of the CES-D by assessing other constructs (e.g., perceived social competence) and symptoms of other disorders (e.yard., Social Anxiety Disorder) [iv], [11], [14], [21]. For at least ane item (i.e., "I had crying spells"), there appears to be a robust sex difference in responses, leading to inappropriate aggrandizement of women'due south CES-D scores due to cultural norms regarding emotional expression, rather than actual differences in depressive symptoms [19], [20], [25], [28], [29]. Furthermore, the CES-D also includes four reverse-worded items (e.thousand., "I was happy") designed "…to break tendencies toward response prepare besides as to appraise positive touch (or its absence)" [2]; however, these ii purposes are at odds and may lead to misrepresentation of response patterns or biased estimations of positive bear on [4], [30]. Research suggests that depression marked past absence of positive affect (i.e., anhedonia) may be qualitatively and quantitatively different than depression resulting from heightened negative touch on [31]–[33], implying that measures of low should assess this dimension direct.
The aims of the current report were to (1) identify any sex activity biases within the particular content of the CES-D, (2) explore which of the many prior factor solutions for the CES-D (Table i) would demonstrate the best factorial validity, and (iii) test whether a new theory-driven solution would exhibit the best fit. The ability of items to predict depression similarly among men and women (i.e., differential validity) was assessed by using an awarding of detail response theory. The factorial validity of the CES-D was examined using a series of confirmatory factor analyses (CFAs) that tested previously established models, besides as new models based on theory and empirical enquiry. This arroyo is in line with conclusions from a contempo meta-analysis [4] suggesting that the apply of CFAs would be an advisable adjacent footstep in solidifying the optimal factor structure of the CES-D; that is, the utilise of CFAs will circumvent the almost exclusive prior use of exploratory gene analytic techniques with the CES-D [4], [xv]. The present study performed these analyses using five dissimilar samples (i.e., undergraduate, customs, rehabilitation, clinical with a history of depression, and a nationally representative sample from the National Health and Nutrition Test Survey; NHANES) to let generalizability of the findings beyond several applications (east.m., epidemiological, clinical), while addressing the overuse of information from specialized samples in this surface area (due east.g., adolescent, geriatric).
Methods
Ethics Argument
The present study has been ethically approved by the University of Regina Research Ethics Board. The study uses archival data from several sources (details beneath); all the same, participants provided written informed consent prior to participating in the data collection associated with each archival source. The consent forms in those data collections were all approved by ethics committees.
Participants
The first sample included undergraduates (n = 948) from the University of Regina (251 men, 18–52 years [Chiliad historic period = 21.2; SD = iv.three] and 697 women, 18–fifty years [M age = 21.0; SD = 4.7]) who completed the CES-D every bit part of other investigations approved by the Academy of Regina Research Ethics Board. Using this type of sample by and large ensures a wide range of responses, whereas an entirely clinical sample might provide a restricted range of relatively college responses [15], [34]. Participants identified their ethnicity as White/Caucasian (89%), First Nations (i.e., Canadian Aboriginal; three%), Asian (iv%), or other (4%). Most reported existence unmarried (84%), while others were married or cohabiting (13%), separated or divorced (1%), or chose not to answer (2%). Undergraduates were recruited via campus advertisements directing them to a secure website for completion of an online questionnaire parcel.
The 2nd sample included customs members (n = 254) from across Canada (73 men, 18–54 years [1000 age = 32.half dozen; SD = 11.3] and 181 women, 18–55 years [1000 historic period = 32.0; SD = 11.3]) who completed the CES-D as part of some other web-based investigation approved past the University of Regina Enquiry Ethics Board. Similar the undergraduate sample, the community sample was included to ensure a wide range of responses. Most (seventy%) reported having at least some postsecondary didactics, being employed (l% full-time, fourteen% part-fourth dimension, ten% as homemakers), and existence single (52%). Others reported being married or cohabiting (35%), separated or divorced (10%), or chose not to reply (three%). Participants identified their ethnicity as Caucasian (87%), First Nations (Canadian Aboriginal; two%), Asian (2%), or other (9%).
The third sample was a rehabilitation sample of 3rd level rehabilitation patients (n = 522) from a regime-sponsored rehabilitation program who completed the CES-D every bit part of tertiary assessment for bug related to injuries sustained in motor-vehicle or piece of work place accidents (246 men, xviii–85 years [M age = 42.5; SD = 12.5] and 276 women, eighteen–79 years [M age = 43.2; SD = 12.five]). The rehabilitation sample was included to provide a comparatively wide range of responses from a treatment-seeking sample that is very probable distressed, but not necessarily depressed. Ethnicity information was not recorded for the rehabilitation sample, but can exist assumed to exist primarily Caucasian based on population demographics. Virtually reported beingness married or cohabiting (57%), while others were unmarried (27%), separated or divorced (thirteen%), or widowed (3%). Education levels were not available for this sample.
The 4th sample, described as a clinical sample, included community members (n = 84) from across Canada (19 men, eighteen–53 years [M historic period = 29.4; SD = 11.4] and 65 women, 18–55 years [1000 age = 24.4; SD = eight.4]) who completed the CES-D as part of some other web-based investigation approved by the University of Regina Research Ideals Lath. In this sample, participants reported being diagnosed with Major Depressive Disorder past a psychiatrist (77%) or a registered doctoral level psychologist (23%). The average reported length of time since diagnosis was approximately iv years. Most of the clinical participants (60%) reported having at least some postsecondary education, and nigh reported being employed (24% full-time, 20% part-time) or students (39%). Clinical participants identified their ethnicity as Caucasian (89%), Showtime Nations (i.e., Canadian Aboriginal; 4%), or other (7%). Participants reported being single (63%), married or cohabiting (29%), or separated or divorced (8%).
The fifth sample, referred to throughout every bit the NHANES sample, included community members (n = 2814) from a big scale sampling of participants across the United States (1242 men, 25–74 years [M age = 46.5; SD = fourteen.0] and 1572 women, 25–74 years [Thousand age = 45.one; SD = 13.9]) who completed the CES-D. The data was collected by the National Eye for Health Statistics from 1971–1975 as function of a Wellness and Diet Examination Survey; however, depression symptoms have not inverse substantially since then [i], [8]. The public access information is from the National Institute of Mental Health and we are grateful for the NHANES contribution. Comprehensive descriptions of the data collection are available straight online from the Centres for Disease Control and Prevention. Many of the NHANES participants reported having completed Course 12 (37%) or having at least some postsecondary education (32%), and most reported being employed (52% full-time, 11% role-time) or working every bit homemakers (33%). NHANES participants identified their ethnicity equally Caucasian (91%), African American (8%), or other (1%). The majority reported being married or cohabiting (79%), while others reported being single (7%), separated or divorced (eight%), or widowed (6%).
Measures
The CES-D is a 20-item measure assessing symptoms of depression with items phrased as self-statements (e.grand., "I felt hopeful about the time to come"). Respondents rate how frequently each particular practical to them over the course of the by week. Ratings were based on a 4-point Likert scale ranging from 0 (rarely or none of the time [less than 1 24-hour interval]) to 3 (about or all of the time [v–7 days]).
Analyses
Descriptive statistics and differential item functioning
Descriptive statistics were calculated for each detail within each of the samples (Table 2). Means on each of the items for men and women were compared by t-tests across samples every bit an initial alphabetize of differential validity. Differential item operation was after estimated to appraise whether men and women differed in their responses to each item forth the continuum of CES-D scores. Differential item functioning occurs when individuals with the same latent trait (i.e., depression) or total score (e.grand., on the CES-D) reply to items differently due to exam characteristics (e.g., paper and pencil vs. computerised) or biases (e.g., due to sex or race [35], [36]). Estimates of differential item performance can illustrate, for example, that men and women may respond similarly to an item when they have relatively low CES-D scores, only answer differently to the item when they are severely depressed. Differential item performance was estimated using an item response theory approach rather than a Mantel-Haenszel arroyo as it provides a more accurate approximate of non-uniform differential detail functioning (eastward.yard., if it occurs only in more severe levels of depression [37]). Non-parametric item characteristic curves were rendered using jMetrik 2.1.0 [38] and were smoothed using a Gaussian kernel. Item feature curves are an integral part of item response theory that plot which response option (east.g., 0, 1, 2, or 3 on a Likert scale) is nigh likely to be endorsed by an individual with a certain total score. To illustrate an absence of differential particular functioning on the CES-D, men and women with similar levels of depression should endorse the same option on each item of the CES-D (e.g., severely depressed men and women would both chose the highest pick), and therefore showroom very similar item characteristic curves. The distance between the curves for each sex activity was examined manually to identify potential differential item performance. An item was only confidently deemed to exhibit differential particular functioning if the curves for men and women were grossly different either in slope or intercept. Item response theory analyses require both relatively large samples and a range of scores spanning the total continuum of potential scores on the measure [35]; consequently, all five samples were combined for these, but not subsequent analyses. Item characteristic curves were plotted based on total CES-D scores, rather than latent depression, given the aforementioned difficulties associated with the latent construction of the CES-D.
Tabular array ii
Descriptive statistics.
| Undergraduate Sample (n = 948) | Community Sample (northward = 254) | Rehabilitation Sample (n = 522) | Clinical Sample (n = 84) | NHANES (n = 2814) | |||||||||||
| M (SD) | South (.08) | Grand (.16) | M(SD) | S (.fifteen) | K (.30) | M (SD) | S (.11) | One thousand (.21) | M(SD) | S (.26) | M (.52) | M (SD) | Southward (.05) | Thousand (.09) | |
| CES-D ane | .58 (.79) | 1.26 | .96 | .54 (.76) | i.20 | .57 | .87 (.94) | .77 | −.44 | 1.27 (.97) | .14 | −1.02 | .34 (.66) | 2.11 | 4.22 |
| CES-D two | .58 (.85) | 1.40 | i.15 | .75 (.96) | 1.02 | −.fourteen | .82 (1.00) | .90 | −.43 | 1.61 (one.16) | −.17 | −i.43 | .29 (.69) | two.56 | five.98 |
| CES-D 3 | .74 (.94) | 1.01 | −.10 | .96 (ane.05) | .75 | −.71 | .68 (.94) | 1.22 | .37 | i.98 (.99) | −.56 | −.81 | .24 (.64) | 2.91 | 8.13 |
| CES-D 4* | 1.05 (1.05) | .55 | −.96 | 1.27 (ane.14) | .25 | −1.38 | .88 (ane.05) | .87 | −.57 | two.25 (.92) | −1.10 | .36 | .73 (1.19) | ane.nineteen | −.38 |
| CES-D five | 1.15 (.92) | .37 | −.71 | i.01 (.95) | .57 | −.66 | i.17 (ane.05) | .40 | −one.06 | 2.00 (one.03) | −.61 | −.85 | .41 (.73) | ane.83 | 2.82 |
| CES-D 6 | .79 (.95) | .95 | −.twenty | 1.22 (i.eleven) | .37 | −i.21 | .77 (.97) | 1.02 | −.11 | 2.40 (.82) | −1.15 | .29 | .44 (.72) | ane.73 | 2.59 |
| CES-D 7 | ane.08 (.94) | .fifty | −.68 | 1.22 (i.10) | .35 | −one.20 | i.55 (ane.08) | −.01 | −1.28 | 2.05 (.99) | −.55 | −.96 | .59 (.92) | 1.51 | 1.xviii |
| CES-D 8* | one.18 (.94) | .37 | −.75 | i.60 (1.eleven) | −.fifteen | −1.31 | .92 (.97) | .70 | −.62 | 2.24 (.79) | −.76 | −.03 | .87 (.91) | −0.91 | −.81 |
| CES-D nine | .xl (.74) | i.xc | ii.93 | .85 (one.06) | .89 | −.59 | .38 (.76) | ii.11 | 3.71 | 1.94 (ane.02) | −.50 | −.94 | .19 (.54) | 3.28 | xi.xv |
| CES-D 10 | .lx (.83) | i.28 | .81 | .85 (1.03) | .ninety | −.46 | .64 (.90) | ane.24 | .52 | 1.60 (1.10) | −.14 | −1.29 | .26 (.61) | 2.65 | seven.06 |
| CES-D 11 | 1.fourteen (one.00) | .45 | −.88 | 1.37 (1.xi) | .18 | −1.31 | 2.00 (1.11) | −.63 | −1.05 | 2.ten (1.05) | −.77 | −.73 | .65 (.89) | 1.23 | .56 |
| CES-D 12* | .97 (.89) | .54 | −.lx | one.40 (1.02) | .10 | −1.09 | .81 (.92) | .86 | −.27 | two.29 (.75) | −.88 | .46 | .60 (.95) | −one.47 | .95 |
| CES-D thirteen | .78 (.85) | .91 | .14 | .93 (.94) | .76 | −.32 | .85 (.98) | .83 | −.51 | 1.38 (1.03) | .12 | −i.11 | .49 (.82) | 1.62 | i.66 |
| CES-D xiv | i.01 (1.00) | .62 | −.75 | one.36 (1.fifteen) | .15 | −i.42 | .72 (.98) | 1.09 | −.05 | ii.46 (.76) | −1.36 | 1.40 | .37 (.74) | 2.09 | iii.69 |
| CES-D 15 | .47 (.71) | ane.49 | i.lxxx | .58 (.85) | 1.34 | .86 | .34 (.69) | ii.24 | 4.86 | 1.20 (one.x) | .32 | −1.25 | .nineteen (.56) | 3.40 | 11.98 |
| CES-D sixteen* | 1.27 (1.07) | .33 | −1.fourteen | 1.twoscore (1.02) | .13 | −ane.08 | .71 (.90) | i.03 | .01 | two.39 (.62) | −.51 | −.61 | .53 (.97) | −1.67 | ane.39 |
| CES-D 17 | .53 (.83) | i.45 | 1.14 | .74 (.99) | i.09 | −.06 | .49 (.86) | ane.68 | 1.75 | 1.40 (1.xiv) | .14 | −1.39 | .16 (.49) | 3.52 | 12.98 |
| CES-D 18 | .96 (.90) | .67 | −.35 | ane.26 (ane.07) | .39 | −1.08 | .82 (.93) | .91 | −.13 | 2.26 (.81) | −.66 | −.72 | .38 (.66) | 1.83 | 3.09 |
| CES-D nineteen | .58 (.83) | 1.37 | 1.06 | .73 (.99) | 1.14 | .07 | .28 (.63) | two.58 | half dozen.80 | ane.86 (1.08) | −.48 | −1.05 | .14 (.47) | 3.76 | fifteen.50 |
| CES-D xx | .91 (.92) | .76 | −.30 | ane.14 (ane.01) | .42 | −.96 | 1.04 (.99) | .58 | −.75 | 2.12 (.91) | −.63 | −.65 | .52 (.78) | i.51 | 1.69 |
Testing and modifying previous factor solutions
A series of CFAs was conducted to replicate and test selected factor structures published in previous studies (Table 1) and to extend these previous models by excluding potentially problematic items as suggested by previous research. Specifically, in that location appears to be consensus throughout the literature that items xv (i.eastward., "People were unfriendly") and xix (i.east., "I felt that people disliked me") may warrant removal as they reflect interpersonal difficulties, a dimension not consequent with contemporary diagnostic criteria for depression [1], [4], [xi], [14], [21]. Similarly, item 17 (i.e., "I had crying spells") may warrant removal as information technology produces robust sex differences in endorsement [25], [28], [29]. Accordingly, previously demonstrated factor structures were tested with and without items 15, 17, and 19. Several previous analyses have as well suggested that two-item factors within the CES-D (Table ane) are inherently unstable [xv], [39]. Given the challenges associated with 2-item factors, models including a 2-item factor (e.k., [21], [40]) were tested with and without the 2-detail factor utilizing the same procedures (i.e., testing with and without items xv, 17, and xix).
CFAs were conducted separately in each sample to determine whether the structure of the CES-D is generalizable and stable across dissimilar applications. The size of the clinical sample was not optimal for CFAs merely inquiry supports the applicability of CFAs in samples of equally low as 51 participants [41]; moreover, the reliability of the factors and the forcefulness of the communalities betwixt the items facilitate the use of CFAs in this sample. The CFAs were performed with AMOS 18 and data from each of the five samples were inputted in a maximum likelihood estimation procedure. Bollen-Stine bootstrap chi-square and computed bootstrapped parameter estimates with estimates from a maximum-likelihood procedure [45], [46] were also conducted considering the data did not showroom multivariate normality; nonetheless, results were comparable to the maximum-likelihood procedure and are excluded for brevity. Each model was evaluated using the following fit indices with xc% confidence intervals (when applicative): i) chi-square (values should not be significant); 2) chi-square/df ratio (values should be less than 2.0); iii) Comparative Fit Alphabetize (CFI; values must be greater than.90, and platonic fits approach or are greater than.95); four) the Standardized Root Mean Square Rest (SRMR; values must be less than.10 and ideal fits approach or are less than.05); 5) Root Hateful Square Mistake of Approximation (RMSEA; values must be less than.08 and ideal fits arroyo or are less than.05, with ninety% confidence interval values below.10); and vi) Expected Cross-Validation Index (ECVI; when comparing these scores across unlike models, lower values indicate a closer fit [42], [43]. Evaluations emphasized the latter four fit indices (i.e., CFI, SRMR, RMSEA, and ECVI) [44]. Given the large number of models that were tested, but fit indices for solutions where the CFI exceeded.92 in at least three of the v samples were included for presentation.
Results
Internal Consistency
Internal consistency was adequate for the current undergraduate (Cronbach'due south α = .91), customs (Cronbach's α = .94), rehabilitation (Cronbach'due south α = .92), clinical (Cronbach's α = .85), and NHANES (Cronbach'due south α = .85) samples. The average inter-item Pearson correlation with the opposite-scored items (i.e., positive affect/anhedonia) was .34 for the undergraduate sample, .43 for the community sample, .38 for the rehabilitation sample, .23 for the clinical sample, and .26 for the NHANES sample. The average inter-item Pearson correlation without the reverse-scored items (i.eastward., positive impact/anhedonia) was .37 for the undergraduate sample, .44 for the community sample, .twoscore for the rehabilitation sample, .25 for the clinical sample, and .33 for the NHANES sample. In all cases the boilerplate inter-particular correlation was relatively low, indicating variety among the items and supporting notions of more than 1 latent construct. The lowest inter-item correlation was for the clinical sample and suggests that there may be substantial variation amidst clinical presentations of these symptoms for persons with a history of depression. Such variation is implicitly supported by DSM-IV-TR diagnostic criteria that allow for high levels of negative affect or high levels of anhedonia to qualify equally hallmark criteria for major depressive disorder (i.e., "(1) depressed mood or (2) loss of interest or pleasance"; page 356 [1]).
Sex activity Differences on CES-D Items
Across all samples, persons with missing data (i.east., fewer than one%) were excluded from the analyses. The t-tests comparison men and women'due south responses from all samples combined suggested that women reported statistically significantly higher scores (p<.05) on most CES-D items (i.e., 1, two, 3, v, half-dozen, ix, 10, eleven, 12, 14, 15, 16, 17, eighteen, 19, xx); however, the effect sizes (i.e., using percentage of variance accounted for "r 2") were negligible (i.due east., r two<.01) for most, but not all items (i.e., items 3, v, 6, 20, r 2 = .02; item fourteen, r 2 = .03; detail 18, r 2 = .04; item 17, r 2 = .07). Detail 17 (i.e., "I had crying spells") was the simply item with detail characteristic curves that differed markedly between men and women, suggesting it has significant differential item operation. An particular with zilch or negligible differential item operation (i.eastward., item 20) is presented in Figure 1 (i.e., Item characteristic curves) alongside item 17 for illustrative purposes. The item feature curves demonstrate that men and women answer similarly to particular 17 when depression levels are depression or slightly higher up average (−2.v SD to +0.5 SD), with both sexes choosing 0 (rarely or none of the time); however, equally depression levels increase, women are more likely to choose a college response option compared to men. Indeed, fifty-fifty the nearly depressed men are nearly likely to cull 1 (some or a little of the time), while the almost depressed women are more probable to choose 2 (occasionally or a moderate amount of the time) or 3 (most or all of the time). The item feature curve plots for all items are not displayed for brevity, but are available from the authors upon request.
Particular feature curves.
Structural Analyses: CFA Results
The fit indices for each of the previously reported models – equally evaluated with data from each sample – are presented in Table 3 (where the model CFI exceeded.92 in at to the lowest degree three out of the five samples). The results were interpreted to suggest that five models might have the factorial validity to provide utility in divergent populations, as many of the fit indices met adequate standards beyond the different samples. Nonetheless, all of these models included item 17 and/or failed to include items that appraise positive bear upon, which is inconsistent with current theory and diagnostic approaches concerning depression [one]. Of all the newly derived models (i.e., with items 15, 17, and 19 removed and without ii-item factors [if relevant]), only ane exhibited adequate fit indices within each sample, included positive affect items, and did non include item 17. The model with the best fit indices was a revision of the i proposed by Radloff [47], which also excluded items nine, ten, and 13. Relevant fit indices and inter-gene correlations for this newly derived model are reported in Table four. The original model proposed by Radloff [47] included four factors: depressed affect (items 3, 6, 14, 17, 18), anhedonia (items 4, 8, 12, 16), somatic complaints (items one, 2, five, 7, 11, 20), and interpersonal concerns (items fifteen, 19). Eliminating item 17 and the ii interpersonal items results in an easily interpretable 3-factor structure (Tables 5 and 6; Effigy 2– Path Diagram for the CES-D new cistron solution) that includes factors of negative affect (items 3, 6, fourteen, 18), anhedonia (items four, eight, 12, xvi), and somatic complaints (items one, 2, 5, seven, xi, twenty), which is compatible with current DSM-Four-TR conceptualization of depression [1]. The internal consistencies (determined using Cronbach'south alpha) for the total score of the newly derived factor structure (undergraduate α = .87; customs α = .92; rehabilitation α = .90; clinical α = .80; NHANES α = .83), the negative bear on subscale (undergraduate α = .87; community α = .90; rehabilitation α = .89; clinical α = .82; NHANES α = .74), the anhedonia subscale (undergraduate α = .75; community α = .86; rehabilitation α = .79; clinical α = .81; NHANES α = .73), and the somatic subscale (undergraduate α = .72; community α = .80; rehabilitation α = .78; clinical α = .51; NHANES α = .81) were all acceptable with the exception of the somatic subscale in the clinical sample (i.eastward., α = .51). The correlation betwixt the full score of the original CES-D and the total score of the electric current variant, besides every bit the correlations betwixt their corresponding subscale scores, were all very high (Table 7).
Path Diagram for the CES-D new factor solution.
Table 3
CFA fit indices of prior models using electric current samples and sorted past publication date.
| Reference | Factors (Items) | Sample | Xii | df | Xtwo/df | CFI | SRMR | RMSEA | RMSEA 90% CI | ECVI | ECVI 90% CI |
| Radloff, 1977 Model B [ii] | 4 (20)† | Undergraduate | 741.50 | 164 | four.52 | .92 | .04 | .06 | .06;.07 | .88 | .80;.97 |
| Community | 335.38 | 164 | 2.05 | .94 | .04 | .06 | .05;.07 | i.69 | 1.50; 1.91 | ||
| Rehabilitation | 501.36 | 164 | three.06 | .93 | .05 | .06 | .06;.07 | 1.13 | 1.02; 1.28 | ||
| Clinical | 218.02 | 164 | 1.33 | .89 | .08 | .06 | .04;.08 | 3.74 | 3.32; 4.25 | ||
| NHANES | 1209.22 | 164 | seven.37 | .94 | .04 | .05 | .05;.05 | .46 | .42;.l | ||
| Shrout, 1989 Model A [58] | 1 (5)‡ † | Undergraduate | 4.76 | five | .95 | 1.00 | .01 | <.01 | <.01;.04 | .03 | .03;.04 |
| Customs | 4.45 | v | .89 | 1.00 | .02 | <.01 | <.01;.08 | .10 | .x;.13 | ||
| Rehabilitation | 24.58 | 5 | 4.92 | .97 | .04 | .09 | .05;.12 | .09 | .06;.12 | ||
| Clinical | 4.54 | 5 | .91 | ane.00 | .05 | <.01 | .00;.15 | .30 | .thirty;.41 | ||
| NHANES | 102.06 | 5 | twenty.41 | .97 | .03 | .08 | .06;.10 | .04 | .03;.06 | ||
| Burnam, 1988 [66]; Tuunainen, 2001 [67] | 1 (6)† | Undergraduate | 74.82 | ix | 8.31 | .95 | .04 | .10 | .08;.12 | .12 | .09;.16 |
| Community | 18.95 | 9 | 2.11 | .99 | .02 | .05 | .02;.09 | .11 | .09;.15 | ||
| Rehabilitation | 22.95 | 9 | 2.55 | .99 | .02 | .06 | .03;.08 | .09 | .07;.13 | ||
| Clinical | 8.73 | 9 | .97 | 1.00 | .05 | .00 | .00;.12 | .39 | .40;.53 | ||
| NHANES | 60.66 | 9 | 6.74 | .99 | .02 | .05 | .04;.06 | .03 | .02;.04 | ||
| Stansbury, 2006 [11] | 3 (16)‡ † | Undergraduate | 557.32 | 101 | five.52 | .92 | .04 | .07 | .06;.08 | .66 | .59;.75 |
| Community | 216.94 | 101 | 2.15 | .94 | .04 | .07 | .06;.08 | 1.13 | .98; 1.32 | ||
| Rehabilitation | 351.30 | 101 | iii.48 | .94 | .05 | .07 | .06;.08 | .81 | .71;.93 | ||
| Clinical | 119.82 | 101 | 1.19 | .94 | .07 | .05 | .00;.08 | 2.29 | 2.06; 2.50 | ||
| NHANES | 833.84 | 101 | viii.26 | .95 | .03 | .05 | .05;.05 | .32 | .29;.36 | ||
| Lee SW, 2008 Model B [xiii] | three (16)‡ † | Undergraduate | 562.94 | 101 | v.57 | .92 | .04 | .07 | .06;.08 | .67 | .59;.75 |
| Community | 261.13 | 101 | 2.59 | .92 | .05 | .08 | .07;.09 | 1.31 | 1.14; 1.51 | ||
| Rehabilitation | 402.09 | 101 | 3.98 | .92 | .05 | .08 | .07;.08 | .91 | .80; 1.03 | ||
| Clinical | 122.00 | 101 | 1.21 | .94 | .07 | .05 | .00;.08 | 2.31 | 2.06; two.70 | ||
| NHANES | 990.73 | 101 | 9.81 | .94 | .04 | .06 | .05;.06 | .38 | .34;.41 |
Table 4
Newly derived 3-factor xiv-detail solution and associated CFA fit indices.
| Inter-factor Correlations | ||||||||||
| Sample | X2 | df | X2/df | CFI | SRMR | RMSEA (90% CI) | ECVI (ninety% CI) | ane∶2 | two∶three | ane∶three |
| Undergraduate | 286.79 | 74 | 3.88 | .96 | .04 | .06 (.05;.06) | .37 (.32;.43) | .79 | .57 | .75 |
| Community | 151.35 | 74 | 2.05 | .96 | .04 | .06 (.05;.08) | .84 (.72; 1.00) | .83 | .66 | .74 |
| Rehabilitation | 174.64 | 74 | ii.36 | .97 | .04 | .05 (.04;.06) | .45 (.39;.54) | .84 | .71 | .eighty |
| Clinical | 87.24 | 74 | 1.18 | .96 | .07 | .05 (<.01;.08) | 1.fourscore (one.64; 2.13) | .85 | .60 | .46 |
| NHANES | 556.11 | 74 | 7.52 | .96 | .04 | .05 (.04;.05) | .22 (.19;.25) | .87 | .35 | .30 |
Tabular array 5
The 14 items from the original CES-D included in the new solution and their assigned factors.
| CES-D Original items | New 3 factors |
| one. I was bothered by things that unremarkably don't carp me. | 3 |
| 2. I did non experience like eating; my appetite was poor. | iii |
| 3. I felt that I could not milk shake off the blues, even with the help from family or friends. | 1 |
| 4. I felt that I was only as good as other people. | ii |
| 5. I had trouble keeping my listen on what I was doing. | three |
| 6. I felt depressed. | 1 |
| seven. I felt that everything I did was an effort. | 3 |
| 8. I felt hopeful about the hereafter. | 2 |
| 9. I idea my life had been a failure. | - |
| x. I felt fearful. | - |
| xi. My slumber was restless. | 3 |
| 12. I was happy. | 2 |
| xiii. I talked less than usual. | - |
| 14. I felt lonely. | ane |
| 15. People were unfriendly. | - |
| 16. I enjoyed life. | 2 |
| 17. I had crying spells. | - |
| xviii. I felt deplorable. | 1 |
| nineteen. I felt that people disliked me. | - |
| 20. I could not get "going". | three |
Tabular array six
Loading weights and residuals for the CES-D new factor solution.
| Undergraduate Sample | Community Sample | Rehabilitation Sample | Clinical Sample | NHANES Sample | ||
| Item Number | Weight (Remainder) | Weight (Residuum) | Weight (Balance) | Weight (Residual) | Weight (Residuum) | |
| Somatic Symptoms | 1 | .58 (.33) | .59 (.35) | .72 (.52) | .56 (.31) | .59 (.35) |
| 2 | .49 (.24) | .57 (.33) | .61 (.37) | .09 (.01) | .51 (.26) | |
| v | .57 (.33) | .63 (.39) | .65 (.43) | .56 (.31) | .64 (.41) | |
| vii | .53 (.28) | .73 (.53) | .56 (.32) | .50 (.25) | .59 (.36) | |
| 11 | .46 (.22) | .59 (.35) | .38 (.14) | .11 (.01) | .59 (.36) | |
| 20 | .65 (.42) | .73 (.53) | .70 (.48) | .58 (.33) | .70 (.49) | |
| Negative Affect | three | .82 (.68) | .88 (.77) | .83 (.69) | .68 (.46) | .74 (.55) |
| vi | .85 (.72) | .88 (.77) | .88 (.77) | .89 (.80) | .84 (.71) | |
| xiv | .70 (.49) | .70 (.49) | .73 (.54) | .61 (.38) | .74 (.55) | |
| 18 | .79 (.63) | .86 (.74) | .85 (.72) | .79 (.36) | .81 (.66) | |
| Anhedonia | 4 | .62 (.39) | .62 (.39) | .54 (.29) | .62 (.39) | .47 (.22) |
| 8 | .67 (.44) | .68 (.46) | .58 (.33) | .59 (.35) | .57 (.33) | |
| 12 | .84 (.70) | .91 (.82) | .85 (.72) | .82 (.67) | .fourscore (.64) | |
| 16 | .52 (.27) | .88 (.78) | .82 (.67) | .87 (.75) | .77 (.60) |
Tabular array 7
Inter-subscale correlations for the new CES-D cistron solution.
| CES-D Total | CES-DR Total | CES-DR Negative Bear upon | CES-DR Anhedonia | |
| Undergraduate Sample | ||||
| CES-DR Total | .98 | – | ||
| CES-DR Negative Affect | .ninety | .89 | – | |
| CES-DR Anhedonia | .72 | .76 | .55 | – |
| CES-DR Somatic | .84 | .85 | .68 | .41 |
| Community Sample | ||||
| CES-DR Total | .98 | – | ||
| CES-DR Negative Affect | .91 | .91 | – | |
| CES-DR Anhedonia | .79 | .82 | .65 | – |
| CES-DR Somatic | .87 | .87 | .71 | .53 |
| Rehabilitation Sample | ||||
| CES-DR Total | .99 | – | ||
| CES-DR Negative Affect | .91 | .89 | – | |
| CES-DR Anhedonia | .76 | .fourscore | .64 | – |
| CES-DR Somatic | .86 | .88 | .68 | .51 |
| Clinical Sample | ||||
| CES-DR Full | .95 | – | ||
| CES-DR Negative Affect | .88 | .87 | – | |
| CES-DR Anhedonia | .60 | .70 | .54 | – |
| CES-DR Somatic | .75 | .80 | .55 | .24† |
| NHANES | ||||
| CES-DR Full | .98 | – | ||
| CES-DR Negative Affect | .83 | .79 | – | |
| CES-DR Anhedonia | .59 | .67 | .23 | – |
| CES-DR Somatic | .81 | .lxxx | .68 | .16 |
Discussion
Despite the popularity of the CES-D, there has been considerable debate regarding the optimal factor structure and item content for the measure (see Table 1). The current study sought to summarize and address these issues past assessing the differential validity of the CES-D and comparing the previously proposed cistron solutions for the CES-D to a novel, theoretically-driven model. The results support a fourteen-particular, 3-factor model that is relatively more than congruent with current diagnostic criteria for depression [ane].
Previous research has highlighted that item 17 (i.east., "I had crying spells") of the CES-D may atomic number 82 to inflated scores for women [19], [20], [25], [28], [29]. As expected, detail 17 exhibited significant differential item functioning, such that even the most depressed men were most probable to choose 1 (some or a little of the fourth dimension) on the Likert scale for that particular, compared to the most depressed women, who were more probable to choose 2 (occasionally or a moderate amount of the time) or 3 (nearly or all of the fourth dimension). This finding underscores the importance of removing item 17 from the CES-D and afterward creating and utilizing new norms for the measure that exercise non include this item. Continued use of item 17 and the associated norms or cut-offs volition lead to notable overestimates of low in women and underestimates of depression in men. Such misrepresentations attributable to sexual activity and cultural biases, rather than truthful differences in depression, may accept significant social and practical healthcare implications. Attempting to control for this sex activity difference by subtracting a value from women's scores (e.thou., 1 point off of the full), or past otherwise adjusting norms for each sex would be inappropriate because sexual practice differences on this item are nonlinear (i.e., women score higher compared to men when both are severely depressed). To illustrate, removing one point from women'southward scores would substantially and inappropriately lower scores of women who are on the lower spectrum of depression (i.e., because item 17 is less biased on the lower end of the spectrum) and would nevertheless overestimate the severity of depression in severely depressed women when compared to men.
Results of the CFAs failed to support CES-D models previously identified past exploratory factor analyses. All models with minimally acceptable fit indices for three out of the five samples included private items or 2-item factors that previous research suggests should not be included in the CES-D [25], [28], or involved extreme reductions in item content that impede the capacity of the CES-D to appraise DSM-IV-TR depressive symptoms [ane]. A modified version of the model proposed by Radloff [47] provided a 3-factor (i.eastward., negative affect, anhedonia, and somatic symptoms), fourteen-item solution that is consequent with contemporary conceptualization of low [1] and demonstrated splendid fit within all samples as indicated by all fit indices. The solution also exhibited adequate internal consistency for all factors within all samples, with the exception of the somatic factor having relatively poor internal consistency within the sample with a history of depression. The differing results for internal consistency suggest that negative touch on and anhedonia may be the most characteristic and consistent symptoms of depression, while somatic symptoms may exist more variable betwixt individuals with a history of depression. The differences may result from somatic symptoms being endorsed for reasons other than depression, such as chronic pain.
Several theoretical and clinical implications follow the present findings. Researchers and clinicians should non utilise item 17 of the CES-D (i.due east., "I had crying spells") or exist careful of its use and interpretation. As the current results illustrate, a women crying is non necessarily a viable index of her depression severity − perhaps owing to culture norms of emotional expression − and a lack of crying in either sexual practice is non a viable index of an absenteeism of depression. Utilizing detail 17 may lead to skewed estimations of depression and invalid cut-offs scores. However, crying is a symptom of emotional distress, and researchers should explore the possibility of creating a new item that assesses frequency of crying without a sex bias. For example, perhaps a relative mensurate of crying (eastward.thou., "I cried much more frequently than I commonly do" or "I felt like crying more than usual") rather than an absolute measure of crying (e.1000., "I cried most of the time") may limit such sex biases. Moreover, the current model is consistent with previous findings suggesting that socially-focused items of the CES-D (i.due east., items 15 and 19) should non be included in the mensurate [4], [eleven], [fourteen], [21]. Finally, the electric current results further support low equally a multidimensional disorder consisting of negative affect, anhedonia, and somatic symptoms [48]–[50].
The review of prior studies on the factor structure of the CES-D highlights the divergent results of previous exploratory factor analyses, none of which were strongly supported by CFAs with the present information. Future studies of the CES-D may benefit more than from conducting further theory-driven confirmatory analyses rather than exploratory analyses. The majority of previously reported factor solutions suggested by previous exploratory gene analyses exhibited poor fit in the current samples. The best plumbing fixtures solution was derived from gimmicky theoretical inquiry and previously established empirical data and exhibited excellent fit in the diverseness of samples used. Accordingly, the version of the CES-D presented herein would likely maintain factorial validity across different settings (e.g., clinical, enquiry). Futurity research on the CES-D would benefit from exploring different forms of validity (e.g., convergent validity, predictive validity) with the item set from the model suggested here. In addition, future research designs should explicitly include comments regarding the influence of sample on cistron construction fit indices – a variable that the current results indicates is of import.
Several limitations of the electric current study provide directions for future inquiry. First, the bulk of participants in the electric current samples were not formally evaluated (e.one thousand., with a structured clinical interview) for clinically significant low and although the diagnostic criteria for depression has changed minimally since data for the NHANES was collected (roughly 37 years agone), potential changes over time with respect to social and cultural attitudes may take resulted in different response rates and patterns than if this information was nerveless today. Future research should appraise the sensitivity and specificity of the proposed detail set with participants categorized as meeting or not coming together DSM-Four criteria for Major Depressive Disorder. 2nd, the inability to clinically classify individuals with or without low also precluded estimation of appropriate cut-off scores for the CES-D. Futurity enquiry may benefit from re-examining cutting-off scores while removing items identified in the electric current paper every bit inappropriate. Such an exam may shed light on discrepancies in recommendations for cut-off scores [51]–[56]. Third, including the reverse-scored items that are straightforwardly worded assessments of positive affect/anhedonia may be creating a psychometric bias as a result of incidental response errors. Such a possibility is relatively less probable than using reverse-worded items, but future enquiry could assess for such a bias by examining the items separately and adding a measure out that is non based entirely in self-written report for convergent and divergent validity. Quaternary, combining all five samples created a large enough sample to produce accurate estimations of differential item operation; however, the combination of differing samples (east.g., clinical, community) may accept introduced unmeasured confounds (e.grand., cultural differences in the NHANES but not in the clinical sample) that may impact differential item functioning. Future inquiry should examine differential item functioning on the CES-D in a multifariousness of large, culturally homogeneous samples. 5th, the current study only provides support for a revised version of the CES-D in a primarily English-speaking sample. Hereafter enquiry should cross-validate this revision using a more than culturally diverse sample and test its compatibility with versions of the CES-D in other languages. 6th, the somatic factor included in the final solution demonstrated adequate fit, but relatively low internal consistency. As such, the somatic items may benefit from further revision as they may currently focus on symptoms that are likewise characteristic of other disorders (e.g., anxiety disorders) or fail to appraise symptoms frequently associated with depression. For case, item 11 (i.e., "My sleep was restless") is too vague to be specifically related to depression and certainly excludes hypersomnia, waking early, and difficulty falling comatose, which are characteristic of low [1]. Additional revisions to CES-D content might also consider including items describing cognitive symptoms of depression (e.1000., thoughts of worthlessness or suicidal ideation) to farther adhere to current diagnostic criteria. It may likewise exist worthwhile for future researchers to consider adopting a differential weighting schema for items in the CES-D, such that items are weighted according to their analytical power. That said, given the increasing availability of culling screening measures (eastward.g., PHQ-9 [57]), coupled with the longstanding psychometric difficulties of the scale, it may be time to begin the process of retiring the CES-D in favor of newer measures that are also freely available for utilise.
The present study addressed pertinent issues associated with CES-D items and precedent factor structures. CFAs performed with several samples (i.eastward., undergraduate, community, rehabilitation, clinical, and NHANES) were interpreted to suggest a novel best fitting model for the CES-D that is psychometrically and theoretically robust, comprising three-factors (i.east., negative affect, anhedonia, somatic symptoms) and fourteen-items relatively more than coinciding with current diagnostic criteria for low [1]. The CES-D items may benefit from additional revision; withal, this alternative solution offers a valid item set, without biases related to social concerns or sex activity, for inquiry and clinical applications.
Funding Statement
The authors accept no back up or funding to written report.
References
1. American Psychiatric Association (2000) Diagnostic and statistical transmission of mental disorders (4th ed., text revision). Washington: American Psychiatic Clan. [Google Scholar]
2. Radloff LS (1977) The CES-D calibration: A self-written report depression calibration for inquiry in the general population. Appl Psychol Meas 1: 385–401. [Google Scholar]
3. Fountoulakis KN, Bech P, Panagiotidis P, Siamouli M, Kantartzis Due south, et al. (2007) Comparing of depressive indices: Reliability, validity, relationship to anxiety and personality and the role of age and life events. J Affect Disord 97: 187–195. [PubMed] [Google Scholar]
4. Shafer AB (2006) Meta-assay of the cistron structures of 4 depression questionnaires: Beck, CES-D, Hamilton, and Zung. J Clin Psychol 62: 123–146. [PubMed] [Google Scholar]
5. Zich JM, Attkisson CC, Greenfield TK (1990) Screening for depression in primary care clinics: The CES-D and the BDI. Int J Psychiatry Med 20: 259–277. [PubMed] [Google Scholar]
6. Beck AT, Steer RA, Brawl R, Ranieri WF (1996) Comparison of Brook Low Inventories-IA and -II in psychiatric outpatients. J Pers Assess 67: 588–597. [PubMed] [Google Scholar]
vii. Beck AT, Ward CH, Mendelson M, Mock J, Erbaugh J (1961) An inventory for measuring depression. Arch Gen Psychiatry iv: 561–571. [PubMed] [Google Scholar]
8. American Psychiatric Association (1968) Diagnostic and statistical manual of mental disorders (2nd ed.). Washington: American Psychiatic Clan. [Google Scholar]
9. Callahan CM, Wolinsky FD (1994) The effect of gender and race on the measurement properties of the CES-D in older adults. Med Intendance 32: 341–356. [PubMed] [Google Scholar]
10. Schroevers MJ, Sanderman R, van Sonderen Eastward, Ranchor AV (2000) The evaluation of the Center for Epidemiologic Studies Low (CES-D) calibration: Depressed and positive affect in cancer patients and health reference subjects. Qual Life Res ix: 1015–1029. [PubMed] [Google Scholar]
11. Stansbury JP, Ried LD, Velozo CA (2006) Unidimensionality and bandwidth in the Centre for Epidemiologic Studies Depression (CES-D) calibration. J Pers Assess 86: 10–22. [PubMed] [Google Scholar]
12. Boisvert JA, McCreary DR, Wright KD, Asmundson GJG (2003) Factorial validity of the Center for Epidemiologic Studies-Depression (CES-D) scale in armed services peacekeepers. Depress Anxiety 17: xix–25. [PubMed] [Google Scholar]
thirteen. Lee SW, Stewart SM, Byrne BM, Wong JPS, Ho SY, et al.. (2008) Cistron structure of the Centre for Epidemiological Studies Depression scale in Hong Kong adolescents. J Pers Assess 90, 175–184. [PubMed] [Google Scholar]
14. Williams CD, Taylor TR, Makambi Yard, Harrell J, Palmer JR, et al. (2007) CES-D 4-factor structure is confirmed, but non invariant, in a large cohort of African American women. Psychiatry Res 150: 173–180. [PubMed] [Google Scholar]
15. Osborne JW, editor (2008) Best practices in quantitative methods. Yard Oaks: Sage Publications Inc. 596 p. [Google Scholar]
xvi. Cheng ST, Chan AC, Fung HH (2006) Factorial structure of a curt version of the Center for Epidemiologic Studies Depression scale. Int J Geriatr Psychiatry 21: 333–336. [PubMed] [Google Scholar]
17. Clara I, Cox BJ, Enns MW (2001) Confirmatory cistron analysis of the Low–Anxiety–Stress Scales in depressed and anxious patients. J Psychopathol Behav Assess 23: 61–67. [Google Scholar]
18. Flor H, Kerns RD, Turk DC (1987) The office of spouse reinforcement, perceived pain, and action levels of chronic pain patients. J Psychosom Res 31: 251–259. [PubMed] [Google Scholar]
19. Freedle R, Kostin I (1997) Predicting black and white differential particular functioning in verbal illustration performance. Intelligence 24: 417–444. [Google Scholar]
20. Hunter JE, Schmidt FL (2000) Racial and gender bias in power and achievement tests: Resolving the apparent paradox. Psychol Public Policy Law 6: 151–158. [Google Scholar]
21. Kohout FJ, Berkman LF, Evans DA, Cornoni-Huntley J (1993) Two shorter forms of the CES-D (Center for Epidemiological Studies Depression) depression symptoms index. J Aging Wellness 5: 179–193. [PubMed] [Google Scholar]
22. Lee AE, Chokkanathan Southward (2008) Factor structure of the 10-item CES-D scale among community dwelling older adults in singapore. Int J Geriatr Psychiatry 23: 592–597. [PubMed] [Google Scholar]
23. Peterson CC, Palermo TM (2004) Parental reinforcement of recurrent hurting: The moderating impact of kid low and anxiety on functional disability. J Pediatr Psychol 29: 331–341. [PubMed] [Google Scholar]
24. Pool JJM, Hiralal Southward, Ostelo RWJG, van der Veer K, Vlaeyen JWS, et al. (2009) The applicability of the Tampa Scale of Kinesiophobia for patients with sub-acute neck pain. Qual Quant 43: 773–780. [Google Scholar]
25. Yang FM, Jones RN (2007) Centre for Epidemiologic Studies-Depression scale (CES-D) item response bias establish with Mantel-Haenszel method was successfully replicated using latent variable modeling. J Clin Epidemiol threescore: 1195–1200. [PMC free article] [PubMed] [Google Scholar]
26. Ohayon MM, Schatzberg AF (2003) Using chronic pain to predict depressive morbidity in the general population. Curvation Gen Psychiatry 60: 39–47. [PubMed] [Google Scholar]
27. Snarski 1000, Scogin F (2006) Assessing low in older adults. In: Qualls SH, Knight BG, editors. Psychotherapy for depression in older adults. Hoboken: John Wiley and Sons. 45–77. [Google Scholar]
28. Novak D, Archuleta M, Benson J, Trunnel Eastward, Yipchuck G (1995) The relationship among diet, exercise, and perimenstrual symptoms. J Am Diet Assoc 95: A56. [Google Scholar]
29. Rivera-Medina CL, Caraballo JN, Rodriguez-Cordero ER, Bernal G, Davila-Marrero E (2010) Factor structure of the CES-D and measurement invariance across gender for depression-income puerto ricans in a probability sample. J Consult Clin Psychol 78: 398–408. [PubMed] [Google Scholar]
30. Urbina S (2004) Essentials of psychological testing. Hoboken: John Wiley & Sons. 326 p. [Google Scholar]
31. Clark LA, Watson D (1991) Tripartite model of anxiety and depression: Psychometric evidence and taxonomic implications. J Abnorm Psychol 100: 316–336. [PubMed] [Google Scholar]
32. Nutt D, Demyttenaere Yard, Janka Z, Aarre T, Bourin M, et al. (2007) The other face of depression, reduced positive affect: The role of catecholamines in causation and cure. J Psychopharmacol 21: 461–471. [PubMed] [Google Scholar]
33. Watson D, Clark LA, Carey 1000 (1988) Positive and negative affectivity and their relation to anxiety and depressive disorders. J Abnorm Psychol 97: 346–353. [PubMed] [Google Scholar]
34. Tabachnick BG, Fidell LS (2007) Using multivariate statistics (5th ed.). Boston: Allyn and Bacon. 980 p. [Google Scholar]
35. Embretson SE, Reise SP (2000) Item response theory for psychologists. Mahwah: L. Erlbaum Associates. 371 p. [Google Scholar]
36. Zumbo BD (2007) Three generations of Differential Detail Performance (DIF) analyses: Considering where it has been, where it is at present, and where it is going. Lang Appraise Q 4: 223–233. [Google Scholar]
37. Hambleton RK, Rogers J (1989) Detecting potentially biased test items: Comparison of IRT area and Mantel-Haenszel methods. Appl Meas Educ iv: 313–334. [Google Scholar]
39. Warner RB (2008) Practical statistics: From bivariate through multivariate techniques. London: Sage. 1101 p.
twoscore. Andresen EM, Malmgren JA, Cater WB, Patrick DL (1994) Screening for low in well older adults: Evaluation of a short form of the CES-D (Center for Epidemiologic Studies Depression scale). Am J Prev Med x: 77–84. [PubMed] [Google Scholar]
41. Iacobucci D (2010) Structural equations modeling: Fit Indices, sample size, and advanced topics. J Consum Psychol twenty: 90–98. [Google Scholar]
42. Browne MW, Cudeck R (1989) Single sample cantankerous-validation indices for covariance structures. Multivariate Behav Res 24: 445–455. [PubMed] [Google Scholar]
43. Browne MW, Cudeck R (1993) Alternative ways of assessing model fit. In: Bollen KA, Long JS, editors. Testing structural equation models. Newbury Park: Sage Publications Inc. 136–162. [Google Scholar]
44. Hu L, Bentler PM (1999) Fit indices in covariance structure modeling: Sensitivity to underparameterized model mis-specification. Psychol Methods 3: 424–453. [Google Scholar]
45. Byrne B (2001) Structural equation modeling with Amos: Bones concepts, applications, and programming. Mahwah: Erlbaum. 352 p.
46. Nevitt J, Hancock GR (2001) Performance of bootstrapping approaches to model test statistics and parameter standard error interpretation in structural equation modeling. Struct Equ Modeling 8: 353–377. [Google Scholar]
47. Radloff LS (1991) The utilise of the Heart for Epidemiologic Studies Depression scale in adolescents and young adults. J Youth Adolescence 20: 149–166. [PubMed] [Google Scholar]
48. Morley S, Williams AC, Black S (2002) A confirmatory gene assay of the Beck Depression Inventory in chronic pain. Pain 99: 289–298. [PubMed] [Google Scholar]
49. Vanheule S, Desmet M, Groenvynck H, Rosseel Y, Fontaine J (2008) The factor structure of the Beck Depression Inventory-Ii: An evaluation. Assessment 15: 177–187. [PubMed] [Google Scholar]
50. Ward LC (2006) Comparison of factor structure models for the Brook Low Inventory-Two. Psychol Assess 18: 81–88. [PubMed] [Google Scholar]
51. Boyd JH, Weissman MM, Thompson WD, Myers JK (1982) Screening for depression in a community sample. Understanding the discrepancies between depression symptom and diagnostic scales. Curvation Gen Psychiatry 39: 1195–1200. [PubMed] [Google Scholar]
52. Roberts RE, Lewinsohn PM, Seeley JR (1991) Screening for boyish depression: A comparing of depression scales. J Am Acad Kid Adolesc Psychiatry 30: 58–66. [PubMed] [Google Scholar]
53. Santor DA, Zuroff DC, Ramsay JO, Cervantes P, Palacios J (1995) Examining scale discriminability in the BDI and CES-D as a function of depressive severity. Psychol Assess 7: 131–139. [Google Scholar]
54. Schulberg HC, Saul M, McClelland M, Ganguli M, Christy W, et al. (1985) Assessing depression in main medical and psychiatric practices. Arch Gen Psychiatry 42: 1164–1170. [PubMed] [Google Scholar]
55. Turk DC, Okifuji A (1994) Detecting low in chronic hurting patients: Capability of self-reports. Behav Res Ther 32: nine–16. [PubMed] [Google Scholar]
56. Weissman MM, Sholomskas D, Pottenger M, Prusoff BA, Locke BZ (1977) Assessing depressive symptoms in v psychiatric populations: A validation study. Am J Epidemiol 106: 203–214. [PubMed] [Google Scholar]
57. Kroenke K, Spitzer RL, Williams JB (2001) The phq-9: Validity of a brief depression severity measure. J Gen Intern Med sixteen: 626–633. [PMC complimentary article] [PubMed] [Google Scholar]
58. Shrout PE, Yager TJ (1989) Reliability and validity of screening scales: Effect of reducing calibration length. J Clin Epidemiol 42: 69–78. [PubMed] [Google Scholar]
59. Carpenter JS, Andrykowski MA, Wilson J, Hall LA, Rayens MK, et al. (1998) Psychometrics for 2 short forms of the Center for Epidemiological Studies - Depression scale. Problems Ment Health Nurs 19: 481–494. [PubMed] [Google Scholar]
60. Irwin M, Artin KH, Oxman MN (1999) Screening for depression in the older adult: Benchmark validity of the 10-item Center for Epidemioligical Studies Low scale (CES-D). Arch Intern Med 159: 1701–1704. [PubMed] [Google Scholar]
61. Santor DA, Coyne JC (1997) Shortening the CES-D to improve its power to discover cases fo low. Psychol Assess 9: 233–243. [Google Scholar]
62. Herrero J, Meneses J (2006) Short spider web-based versions of the Perceived Stress (PSS) and Heart for Epidemiological Studies - Low (CESD) scales: A comparison to pencil and newspaper responses among Net users. Comput Man Behav 22: 830–846. [Google Scholar]
63. Boey KW (1999) Cross-validation of a short form of the CES-D in Chinese Elderly. Int J Geriatr Psychiatry xiv: 608–617. [PubMed] [Google Scholar]
64. Rouch-Leroyer I, Sourgen C, Barberger-Gateau P, Fuhrer R, Dartiques JF (2000) Detection of depressive symptomatology in elderly people: A short version of the CES-D scale. Aging 12: 228–233. [PubMed] [Google Scholar]
65. Yen S, Robins CJ, Lin Due north (2000) A cross-cultural comparison of depressive symptom manifestation: China and the United states of america. J Consult Clin Psychol 68: 993–999. [PubMed] [Google Scholar]
66. Burnam MA, Wells KB, Leake B, Landsverk J (1988) Development of a brief screening musical instrument for detecting depressive disorders. Med Care 26: 775–789. [PubMed] [Google Scholar]
67. Tuunainen A, Langer RD, Klauber MR, Kripke DF (2001) Short version of the CES-D (Burnam screen) for low in reference to the structured psychiatric interview. Psychiatry Res 103: 261–270. [PubMed] [Google Scholar]
68. Bush BA, Novack TA, Schneider JJ, Madan A (2004) Low post-obit traumatic brain injury: The validity of the CES-D every bit a cursory screening device. J Clin Psychol Med Settings eleven: 195–201. [Google Scholar]
69. Cole JC, Rabin As, Smith TL, Kaufman Equally (2004) Development and validation of a Rasch-derived CES-D short form. Psychol Assess sixteen: 360–372. [PubMed] [Google Scholar]
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