Estructura factorial y propiedades psicometricas de la Escala de Actitudes Disfuncionales Revisada en universitarios colombianos.
Dysfunctional schemata are extremely inflexible beliefs that are the main cognitive vulnerability to depression according to the cognitive model advocated by Beck, Rush, Shaw, and Emery (1979). They are thought to be shaped by early negative life experiences, to be relatively stable, and to remain latent until the individual encounters negative events that activate them. In this case, dysfunctional schemata would skew the information processing system, leading to the production of negative automatic thoughts that constitute the cognitive triad (i.e., negative views about oneself, the world, and the future).
The measurement of dysfunctional schemata has been mainly conducted by applying the Dysfunctional Attitude Scale (DAS; Weissman & Beck, 1978). Most of the studies have relied on the total score of the DAS as a general cognitive vulnerability to depression, as exploratory factor analyses (EFA) have yielded mixed results regarding the number of factors extracted, with studies finding between two- to four-factor solutions (e.g., Cane, Olinger, Gotlib, & Kuiper, 1986; Chioqueta & Stiles, 2006; Sanz & Vazquez, 1993). Accordingly, de Graaf, Roelofs, and Huibers (2009) used confirmatory factor analysis (CFA) to compare the fit of the previously proposed factor structures using a Dutch version of the DAS with a very large general population sample (N = 8960). The authors found that the two-factor solution was the most adequate fit of the data and revised the DAS by retaining 17 items of the 40 original ones. This revised version (hereafter the DAS-R) consists of two correlated factors labeled as perfectionism/performance evaluation and dependency.
De Graaf et al. (2009) recommended the use of the DASR because it has some advantages over the full version. First, the DAS-R showed a clearer factor structure than the DAS and possesses good psychometric properties in terms of model fit, reliability, and convergent construct validity. Second, the DAS-R can considerably shorten the administration time with respect to the full DAS scale. Lastly, the DAS-R contains two theoretically meaningful subscales that measure specific dysfunctional schemata. This constitutes an advance in the analysis of the cognitive model of depression because, according to Beck (1987), vulnerable individuals might show only specific dysfunctional schemata rather than the whole range of dysfunctional beliefs measured by the DAS.
Following the work by de Graaf, Roelofs, & Huibers (2009), Ruiz et al. (2015) analyzed the factor structure and psychometric properties of the DAS-R in a Spanish sample mostly formed by undergraduates using the DAS version by Sanz and Vazquez (1993). The DAS-R showed excellent internal consistency and discriminant and convergent validity. The same two-factor structure as in de Graaf et al.'s study was found. Further, Ruiz et al. provided evidence of a hierarchical structure with two first-order factors (perfectionism/performance evaluation and dependency) and a second-order factor that reflects dysfunctional schemata in general. This finding is particularly important because it supports the common practice of aggregating DAS items into only one score versus calculating the subscales scores. This way, researchers and clinicians have more flexibility because they can obtain a global score of the DAS-R or separate scores of its two first-order factors depending on their interests.
To our best knowledge, neither the DAS nor the DAS-R have been validated in Colombia. The current study aimed at analyzing the psychometric properties and factor structure of the DAS-R by Ruiz et al. (2015) in a Colombian sample of undergraduates (N = 762).
The sample included 762 undergraduates (age range 18-63, M = 21.16, SD = 3.76) from four universities of Bogota. Forty-six percent of the sample were Psychology undergraduate students. The other majors included Law, Engineering, Mathematics, and Physics. Sixty-two percent were women. Of the overall sample, 26% of participants had received psychological or psychiatric treatment at some time, but only 4.3% were currently in treatment. Also, 2.9% of participants were taking some psychotropic medication.
Dysfunctional Attitude Scale--Revised
(DAS-R; de Graaf et al., 2009; Weissman & Beck, 1978; Spanish version by Ruiz et al., 2015). The DAS comprises 40 items that are rated on a 7-point Likert-type scale (7 = fully agree; 1 = fully disagree). The revised version of the DAS (i.e., the DAS-R; de Graaf et al., 2009) contains 17 items grouped in two factors: perfectionism/performance evaluation (e.g., "It is difficult to be happy unless one is good-looking, intelligent, rich and creative") and dependency (e.g., "My value as a person depends greatly on what others think of me"). The Spanish version of the DAS-R showed excellent internal consistency for the total scale ([alpha] = .90), and good internal consistency for the perfectionism/performance evaluation factor ([alpha] = .87) and the dependency factor ([alpha] = .81). A factor structure with two-correlated factors and a second-order factor was obtained.
Automatic Thoughts Questionnaire--8
(ATQ-8; Netemeyer et al., 2002; Spanish version by Cano-Garcia & Rodriguez-Franco, 2002). The ATQis a measure of the frequency of negative automatic thoughts experienced during the past week. It consists of 8 negative automatic thoughts that are rated on a 5-point Likert-type scale (5 = all the time; 1 = not at all). Examples of items are "I'm no good," "Nothing feels good anymore," "What's wrong with me?" and "I'm worthless." The ATQ-8 showed good internal consistency in this study ([alpha] = .85). According to the cognitive theory of depression, it was expected that the DAS-R would show medium to strong correlations with the ATQ-8.
Acceptance and Action Questionnaire--11
(AAQ-II; Bond et al., 2011; Spanish version by Ruiz, Langer, Luciano, Cangas, & Beltran, 2013). The AAQ-II is a general measure of psychological inflexibility. It consists of 7 items that are rated on a 7-point Likert-type scale (7 = always true; 1 = never true). The items reflect unwillingness to experience unwanted emotions and thoughts (e.g., "I worry about not being able to control my worries and feelings") and the inability to be in the present moment and behave according to value-directed actions when experiencing psychological events that could undermine them (e.g., "My painful experiences and memories make it difficult for me to live a life that I would value"). The alpha found for the AAQ-II in this study was .88. The AAQ-II was administered because psychological inflexibility strongly correlated with the DAS in previous studies (e.g., Ruiz & Odriozola-Gonzalez, in press).
General Health Questionnaire--12
(GHQ-12; Goldberg & Williams, 1988; Spanish version by Rocha, Perez, Rodriguez-Sanz, Borrell, & Obiols, 2011). The GHQ-12 is a 12-item, 4-point Likert-type scale that is frequently used as screening for psychological disorders. Respondents are asked to indicate the degree to which they have recently experienced a range of common symptoms of distress, with higher scores reflecting greater levels of psychological distress. The Likert scoring method was used in this study, with scores ranging 0 to 3 assigned to each of the four response options. The alpha value for the GHQ-12 in this study was .88. Medium to strong correlations were expected between the DAS-R and the GHQ-12.
Depression Anxiety and Stress Scales--21
(DASS-21; Antony, Bieling, Cox, Enns, &Swinson, 1998; Spanish version by Daza, Novy, Stanley, & Averill, 2002). The DASS-21 is a 21-item, 4-point Likert-type scale (3 = applied to me very much, or most of the time; 0 = did not apply to me at all) consisting of sentences describing negative emotional states experienced during the last week. It contains three subscales (depression, anxiety, and stress) and has shown good internal consistency and convergent and discriminant validity. Alpha values in this study were .86, .80, and .80 for the depression, anxiety, and stress subscales, respectively. Medium to strong correlations were expected between the DAS-R and the total score of the DASS-21 and its subscales.
Satisfaction with Life Survey
(SWLS; Diener, Emmons, Larsen, & Griffin, 1985; Spanish version by Atienza, Pons, Balaguer, & Garcia-Merita, 2000). The SWLS is a 5-item, 7-point Likert-type scale (7 = strongly agree; 1 = strongly disagree) that measures self-perceived well-being. Example of items are "I am satisfied with my life" and "In most ways, my life is close to my ideal." The SWLS has good psychometric properties and convergent validity. The alpha value for the SWLS in this study was .85. Medium correlations were expected between the DAS-R and the SWLS.
Following the suggestions by Elosua, Mujika, Almeida, and Hermosilla (2014), a small pilot study was conducted first to explore whether Colombian people experienced difficulties in understanding the items of the Spanish versions of the DAS-R, ATQ-8, AAQ-II, GHQ-12, DASS-21, and SWLS. Ten Colombian undergraduates did not find any difficulties in understanding the DAS-R items; therefore we decided to apply the original scale without changes.
Administration of the questionnaire package was collective and conducted in the participants' classrooms during the beginning of a regular class. Six people administered the questionnaire package following the same instructions. The study was presented, and individuals who signed an informed consent were given a questionnaire packet including the self-report instruments listed above. Upon completion of the study, the participants were debriefed about the aims of the study and thanked for their participation.
Prior to conducting factor analyses, data were examined searching for missing values. Only three values of the DAS-R were missing (one value for items 3, 6, and 14). These data were inputted using the replacing option of the Factor 9.2(c) (Lorenzo-Seva & Ferrando, 2006).
The robustness of the two-factor model with a second-order factor found by Ruiz et al. (2015) and the alternative two-correlated-factor and one-factor models was assessed by conducting confirmatory factor analyses (CFA) using LISREL(c) (version 8.71, Joreskog & Sorbom, 1999) and adopting an unweighted least square estimation method. Goodness-of-fit was examined computing the following fit indexes: (a) the root mean square error of approximation (RMSEA); (b) the comparative fit index (CFI); (c) the non-normed fit index (NNFI); and (d) the expected cross-validation index (ECVI). According to Kelloway (1998), RMSEA values of .10 represent a good fit, with values below .05 representing a very good fit to the data. Regarding the CFI and NNFI, values above .90 indicate well-fitting models, and above .95 represent a very good fit to the data. The ECVI was computed to compare the goodness-of-fit of the three factor structure alternatives (lower values indicate better fit to the model). Lastly, the differences between the chi-square-values for the three models were calculated following a likelihood ratio test under the null hypothesis that the one-factor model fits as well as the two-factor models, and that the two-correlated factor model fits as well as the hierarchical factor model. These chi-square differences are also chi-square distributed with degrees of freedom equal to the difference between the degrees of freedom of the two compared models.
Following the recommendations by Gignac (2007), the Schmid-Leiman transformation (Schmid & Leiman, 1957) was conducted as an alternative to the nested factors modeling to explore the factor loadings of the items and the extracted variance accounted for by the general factor. This procedure performs a secondary EFA using the latent factor intercorrelations obtained from a previous EFA and facilitates interpretation of primary factors (items) relative to higher-order factors by computing direct relations between primary variables and second-order factors. Likewise, the proportion the general factor accounting for the extracted variance is indicative of the presence of a general factor (range = 40-50%; Gorsuch, 1983). This analysis was computed using Factor 9.2(c). An exploratory unweighted least squares factor analysis with direct oblimin rotation and the Schmid-Leiman transformation (Schmid & Leiman, 1957) was conducted. Additionally, the syntax developed by Wolf and Preising (2005) for SPSS was used to compute the total extracted variance accounted for by the higher order factor.
The remaining statistical analyses were performed on SPSS 19(c). Cronbach's alphas providing confidence intervals according to Duhacheck and lacobucci (2004) were computed to explore the internal consistency of the DAS-R. Corrected item-total correlations were obtained to identify items that should be removed because of low discrimination item index (i.e., values below .20). Descriptive data were also calculated. To examine discriminant construct validity, scores on the DAS-R were compared, computing Student's t, between participants with scores above and below the cutoff on the GHQ-12 (i.e., 12 points). Pearson correlations between the DAS-R and the other scales were calculated to assess convergent and divergent construct validity.
Table 1 presents the results of the CFA comparing the three alternative models: (a) one-factor model, (b) two-correlated-factor model, (c) two-factor with a second-order factor model. The one-factor model obtained an acceptable fit, but inferior to the one observed for the two-factor model. The chi-square difference between the two-factor model and the two-factor model with a general factor was 173.53 (df=1, p< .05), thereby indicating that the hierarchical factor model showed a significantly better fit to the data.
Scores on the goodness-of-fit indexes for the hierarchical factor model were good for the RMSEA (RMSEA =.059, 90% Cl [.053, .065]), and very good for the CFI and NNFI (.99 and .98, respectively). Both factors were strongly correlated (r= .83). Table 2 shows the original items, their translation into Spanish, and factor loadings for the two-factor model with a general factor.
According to the Schimd-Leiman transformation, all items of the DAS-R seemed to represent the general factor because they showed loadings above .30 (Tabachnick & Fidell, 2007). The range of factor loadings was between .43 (item 6) and .70 (item 12). The loading of the two first-order factors on the second-order factor were .91 and .92 for the perfectionism/performance evaluation and dependence, respectively. The general factor accounted for 72.4% of the extracted variance, a proportion clearly above the range considered as indicative of the presence of a general factor (40%-50%; Gorsuch, 1983), whereas the two first-order factors explained 22.5% (perfectionism/performance evaluation) and 5.1.% (dependence) of the variance.
Internal Consistency, Descriptive Data and Criterion Validity
Table 3 shows that Cronbach's alpha of the overall DAS-R was .91 (95% Cl [.90, .92]). With respect to the DAS-R factors, Perfectionism/Performance Evaluation showed an alpha of .87 (95% Cl [.86, .89]), whereas the alpha of Dependency was .81 (95% Cl [.79, .83]). Corrected item-total correlations of the DAS-R ranged from .46 to .66. With respect to the two factors, perfectionism/performance evaluation showed item-total correlations between .49 and .68, whereas for dependency, they were between .42 and .65.
Table 4 shows that participants with scores above the cutoff on the GHQ-12 scored statistically significantly higher on the DAS-R and its subscales than those with scores below this cutoff.
Pearson Correlations with other related Constructs
Table 5 shows that the DAS-R showed correlations with all other assessed constructs in theoretically coherent ways. Specifically, the DAS-R showed positive correlations with psychological distress, depression and anxiety symptoms, negative automatic thoughts, and psychological inflexibility; and negative correlations with life satisfaction.
The data obtained in this study provide promising evidence that the DAS-R is a valid and reliable measure of dysfunctional schemata in Colombian samples. Overall, the current data are very similar to those obtained by Ruiz et al. (2015). Specifically, the DAS-R showed excellent internal consistency ([alpha] = .91), with good Cronbach's alphas for its factors (perfectionism/performance evaluation: [alpha] = .87; dependency: [alpha] = .81). The construct convergent validity of the DAS-R was examined by analyzing its correlations with related constructs such as negative automatic thoughts and psychological inflexibility, whereas construct divergent validity was assessed by analyzing DAS-R correlations with life satisfaction. All correlations found were in the expected direction. Although correlations were relatively small, they were similar to those obtained in Ruiz et al. (2015). The DAS-R also presented discriminant validity to the extent that participants who scored above the cutoff on the GHQ-12 scored significantly higher on the DAS-R and its subscales than those who scored below the cutoff.
Importantly, the CFA conducted replicated the hierarchical factor model found by Ruiz et al. (2015) with two-correlated first-order factors (perfectionism/performance evaluation and dependency) and a second-order factor reflecting general dysfunctional schemata. This hierarchical factor structure obtained better fit to the data than the alternative two-correlated factor structure and one-factor structure. As commented in Ruiz et al., this finding has several relevant implications. On the one hand, the presence of a general factor provides a theoretical justification of using the total score of the DAS-R. This score provides a general measure of dysfunctional schemata and not the mere aggregation of the two types of dysfunctional schemata identified. On the other hand, in some contexts, it may be more advisable to analyze the scores on first-order factors. As previously discussed, the possibility of analyzing the presence of specific dysfunctional schemata can be seen as an advance in the study of depression according to cognitive therapy (Beck, 1987).
Some limitations of this study are worth mentioning. Firstly, the functioning of the DAS-R was tested only in a nonclinical sample; therefore, further research is necessary in clinical samples to confirm the results obtained in this study. Secondly, no information was obtained concerning the diagnosis and the course of therapy in participants who reported being in psychological/psychiatric treatment. Thirdly, the sample used in this study consisted of undergraduate individuals and with a narrow age range. Accordingly, further study should analyze the psychometric properties and factor structure of the DAS-R with older people with less education. Fourthly, because all data were obtained using self-report measures, relationships among variables might be artificially inflated. Lastly, the instruments used to explore the convergent validity of the DAS-R lacked of a formal validation in a Colombian sample; however, their internal consistencies were adequate and similar to the ones obtained in the validation studies.
In conclusion, the DAS-R seems to be a reliable and valid measure of dysfunctional schemata in Colombian samples, consisting of a hierarchical factor structure with a general factor and two first-order factors. The DAS-R provides researchers and clinicians the option to investigate specific types of dysfunctional schemata reliably and provides a theoretically meaningful reason for the use of the total score as a general measure of dysfunctional thinking.
Antony, M. M., Bieling, P. J., Cox, B. J., Enns, M. W., & Swinson, R. P. (1998). Psychometric properties of the 42-item and 21-item versions of the Depression Anxiety Stress Scales (DASS) in clinical groups and a community sample. Psychological Assessment, 10, 176-181. http://dx.doi.org/10.1037/1040-35184.108.40.206
Atienza, F. L., Pons, D., Balaguer, I., & Garcia-Merita, M. (2000). Propiedades psicometricas de la Escala de Satisfaccion con la Vida en adolescentes [Psychometric properties of the Satisfaction with Life Scale in adolescents]. Psicothema, 12, 314-319.
Beck, A. T. (1987). Cognitive models of depression. Journal of Cognitive Psychotherapy: An International Quarterly, 1, 5-37.
Beck, A. T., Rush, A. J., Shaw, B. F., & Emery, G. (1979). Cognitive therapy of depression. New York, NY: Guilford.
Bond, F. W., Hayes, S. C., Baer, R. A., Carpenter, K. M., Guenole, N., Orcutt, H. K., et al. (2011). Preliminary psychometric properties of the Acceptance and Action Questionnaire--II: A revised measure of psychological inflexibility and experiential avoidance. Behavior Therapy, 42, 676-688. http://dx.doi.org/10.1016/ j.beth.2011.03.007
Cane, D. B., Olinger, J., Gotlib, I. H., & Kuiper, N. A. (1986). Factor structure of the Dysfunctional Attitude Scale in a student population. Journal of Clinical Psychology, 42, 307-309. http://dx.doi.org/10.1002/1097-4679(198603)42:2<307::AIDJCLP2270420213>3.0.CO;2-J
Cano-Garcia, F. J., & Rodriguez-Franco, L. (2002). Evaluacion del lenguaje interno ansiogeno y depresogeno en la experiencia de dolor cronico [Assessment of anxious and depressive self-talk in chronic pain experience]. Apuntes de Psicologia, 20, 329-346.
Chioqueta, A. R, & Stiles, T. C. (2006). Factor structure of the Dysfunctional Attitude Scale (Form A) and the Automatic Thoughts Questionnaire: An exploratory study. Psychometric Reports, 99, 239-247.
Daza, P., Novy, D. M., Stanley, M., & Averill, P. (2002). The Depression Anxiety Stress Scale-21: Spanish translation and validation with a Hispanic sample. Journal of Psychopathology and Behavioral Assessment, 24, 195-205. http://dx.doi.org/10.1023M: 1016014818163
de Graaf, L. E., Roelofs, J., & Huibers, M. J. H. (2009). Measuring dysfunctional attitudes in the general population: The Dysfunctional Attitude Scale (form A) Revised. Cognitive Therapy and Research, 33, 345-355. http://dx.doi.org/10.1007/s10608-009-9229-y
Diener, E., Emmons, R. A., Larsen, R. J., & Griffin, S. (1985). The Satisfaction with Life Scale. Journal of Personality Assessment, 49, 71-75. http://dx.doi.org/10.1207/s15327752jpa4901_13
Elosua, R, Mujika, J., Almeida, L. S., & Hermosilla, D. (2014). Procedimientos analitico-racionales en la adaptacion de tests. Adaptacion al espanol de la bateria de pruebas de razonamiento [Judgmental-analytical procedures for adapting tests: Adaptation to Spanish of the Reasoning Tests Battery]. Revista Latinoamericana de Psicologia, 46, 117-126. http://dx.doi.org/10.1016/S0120-0534(14)70015-9
Duhacheck, A., & lacobucci, D. (2004). Alpha's standard error (ASE): An accurate and precise confidence interval estimate. Journal of Applied Psychology, 89, 792-808. http://dx.doi.org/10.1037/ 0021-9010.89.5.792
Gignac, G. E. (2007). Multi-factor modeling in individual differences research: Some recommendations and suggestions. Personality and Individual Differences, 42, 37-48. http://dx.doi.org/10.1016/j.paid.2006.06.019
Goldberg, D., & Williams, R (1988). A user's guide to the General Health Questionnaire. Windsor, UK: NFER-Nelson.
Gorsuch, R. L. (1983). Factor analysis (2nd ed.). Hillsdale, NJ: Erlbaum.
Joreskog, K. G., & Sorbom, D. (1999). LISREL 8.30. Chicago, IL: Scientific Software International.
Kelloway, E. K. (1998). Using LISREL for structural equation modeling: A researcher's guide. Thousand Oaks, CA: Sage.
Lorenzo-Seva, U., & Ferrando, R (2006). FACTOR: A computer program to fit the exploratory factor analysis model. Behavior Research Methods, 38, 88-91. http://dx.doi.org/10.3758/ BF03192753
Netemeyer, R. G., Williamson, D. A., Burton, S., Biswas, D., Jindal, S., Landreth, S., et al. (2002). Psychometric properties of shortened versions of the Automatic Thoughts Questionnaire. Educational and Psychological Measurement, 62, 111-129. http://dx.doi.org/10.1177/0013164402062001008
Rocha, K., Perez, K., Rodriguez-Sanz, M., Borrell, C., & Obiols, J. E. (2011). Propiedades psicometricas y valores normativos del General Health Questionnaire (GHQ-12) en poblacion espanola [Psychometric propertie and normative scores of the General Health Questionnaire (GHQ-12) in general Spanish population]. International Journal of Clinical and Health Psychology, 11, 125-139.
Ruiz, F. J., Langer, A. I., Luciano, C., Cangas, A. J., & Beltran, I. (2013). Measuring experiential avoidance and psychological inflexibility: The Spanish translation of the Acceptance and Action Questionnaire. Psicothema, 25, 123-129. http://dx.doi.org/10.7334/psicothema2011.239
Ruiz, F.J., & Odriozola-Gonzalez, P. (in press). The role of psychological inflexibility in Beck's cognitive model of depression in a sample of undergraduates. Anales de Psicologia.
Ruiz, F. J., Suarez-Falcon, J. C., Odriozola-Gonzalez, R, Barbero-Rubio, A., Eisenbeck, N., Lopez-Lopez, J. C., et al. (2015). Factor structure and psychometric properties of the Spanish version of the Dysfunctional Attitude Scale Revised. Behavioral Psychology, 23, 287-303.
Sanz, J., & Vazquez, C. (1993). Adaptacion espanola de la Escala de Actitudes Disfuncionales (DAS) de Weissman and Beck: Propiedades psicometricas y clinicas [Weissman and Beck's Dysfunctional Attitudes Scale Spanish adaptation: Psychometric and clinical properties]. Analisis y Modificacion de Conducta, 19, 705-750.
Schmid, J., & Leiman, J. N. (1957). The development of hierarchical factor solutions. Psychometrika, 22, 53-61. http://dx.doi.org/ 10.1007/BF02289209
Tabachnick, B. G., & Fidell, L. S. (2007). Using multivariate statistics. Boston, MA: Allyn and Bacon.
Weissman, A.N., & Beck, A.T. (1978, November). Development and validation of the Dysfunctional Attitude Scale: A preliminary investigation. Paper presented at the Annual Meeting of the American Educational Research Association, Toronto, Canada.
Wolf, H. G., & Preising, K. (2005). Exploring item and higher order factor structure with the Schmid-Leiman solution: Syntax codes for SPSS and SAS. Behavior Research Methods, 37, 48-58. http://dx.doi.org/10.3758/BF03206397
Francisco J. Ruiz (a) **, Juan Carlos Suarez-Falcon (b), Diego Baron-Rincon (a), Andrea Barrera-Acevedo (a), Alejandra Martinez-Sanchez (a), Andres Pena (a)
(a) Facultad de Psicologia, Fundacion Universitaria Konrad Lorenz, Bogota, Colombia
(b) Facultad de Psicologia, Universidad Nacional de Educacion a Distancia (UNED)
Received 29 July 2015; accepted 16 October 2015
Available online 5 March 2016
* Best Article of the Issue Award
** Corresponding author.
E-mail address: email@example.com (FJ. Ruiz).
Table 1 Goodness-of-Fit indexes of the One-Factor, Two-Correlated Factors, and Two-Correlated Factors with a Second-Order Factor Models of the DAS-R. Goodness-of-fit Two-factor model Two-factor model indicators with a general factor RMSEA [90% CI] .059 [.053, .065] .073 [.067, .079] CFI .99 .98 NNFI .98 .98 ECVI [90% CI] .65 [.58, .74] .88 [.78, .98] [chi square] (df) 425.64 (117) 599.17 (118) Satorra-Bentler Goodness-of-fit One-factor model indicators RMSEA [90% CI] .091 [.085, .096] CFI .97 NNFI .96 ECVI [90% CI] 1.23 [1.11, 1.35] [chi square] (df) 864.68 (119) Satorra-Bentler Table 2 Item Description and Their Factor Loadings in a Completely Standardized Solution. Item number and description Factor loading Perfectionism/performance evaluation 1. Es dificil ser feliz si no se es atractivo, .58 inteligente, rico y creativo [It is difficult to be happy, unless one is good-looking, intelligent, rich, and creative]. 2. Si no hago siempre las cosas bien, la gente no .67 me respetara [If I do not do well all the time, people will not respect me]. 3. Si una persona pide ayuda, es senal de .66 debilidad [If a person asks for help, it is a sign of weakness]. 4. Si no hago las cosas tan bien como los demas, .77 eso significa que soy una persona inferior [If I do not do as well as other people, it means I am an inferior human being] 5. Si fracaso en mi trabajo sere un fracaso como .79 persona [If I fail at my work, then I am a failure as a person]. 6. Si no puedo hacer bien una cosa, es mejor no .57 hacerla. [If you cannot do something well, there is little point in doing it at all]. 7. Si alguien no esta de acuerdo conmigo, eso .66 probablemente indica que no le agrado [If someone disagrees with me, it probably indicates that he does not like me]. 8. Si fracaso en parte, eso lo considero tan malo .78 como ser un completo fracaso [If I fail partly, it is as bad as a complete failure]. 9. Si los demas saben como eres realmente, te .78 consideraran menos [If other people know what you're really like, they will think less of you]. 10. Para ser una persona valiosa debo destacar de .64 verdad por lo menos en un aspecto importante [If I am to be a worthwhile person, I must be truly outstanding in at least one major respect]. 11. Hacer una pregunta me hace parecer inferior .75 [If I ask a question, it makes me look inferior]. Dependency 12. Mi valor como persona depende en gran medida .82 de lo que los demas opinen de mi [My value as a person depends greatly on what others think of me]. 13. Es horrible recibir la censura de personas .53 importantes para uno [It is awful to be disapproved of by people important to you]. 14. Si uno no tiene otras personas en las que .76 confiar, esta destinado a estar triste [If you don't have other people to lean on, you are bound to be sad]. 15. Si desagradas a los demas no puedes ser feliz .77 [If others dislike you, you cannot be happy]. 16. Mi felicidad depende mas de los demas que de .81 mi [My happiness depends more on other people than it does on me]. 17. Es muy importante lo que otras personas .66 piensan sobre mi [What other people think about me is very important]. Table 3 Cronbach's Alphas and Descriptive Data of the Dysfunctional Attitude Scale-Revised. Dysfunctional Attitude Scale-Revised Sample N = 762 (DAS-R) Total Cronbach's alpha .91 [.90, .92] M (SD) 38.27 (16.55) Perfectionism/Performance evaluation Cronbach's alpha .87 [.86, .89] M (SD) 24.53 (11.15) Dependency Cronbach's alpha .81 [.79, .83] M (SD) 13.73 (6.69) Table 4 Mean DAS-R Scores of Participants who Scored above and below the Cutoff of the GHQ-12. Mean SD N DAS-R total score Participants GHQ> 12 43.99 18.48 298 Participants GHQ< 12 34.52 13.99 461 Student's T 7.55 * DAS-R Perfectionism Participants GHQ> 12 28.21 12.22 298 Participants GHQ< 12 22.11 9.68 461 Student's T 7.28 * DAS-R Dependency Participants GHQ> 12 15.77 12.40 298 Participants GHQ< 12 12.40 5.70 461 Student's T 6.57 * * p < .001. Table 5 Pearson Correlations between the DAS-R Scores and Other Self-report Measures. Measure DAS-R-- DAS-R-- DAS-R-- Total Performance Dependency evaluation General Health .31 * .29 * .29 * Questionnaire--12 DASS-21--Depression .42 * .41 * .36 * DASS-21--Anxiety .34 * .33 * .29 * DASS-21--Stress .31 * .29 * .28 * Automatic Thoughts .43 * .41 * .39 * Questionnaire-8 (ATQ-8) Acceptance and Action .42 * .41 * .37 * Questionnaire--II Satisfaction with Life -.26 * -.25 * -.24 * Survey Note. DAS-R: Dysfunctional Attitude Scale-Revised; DASS-21: Depression, Anxiety and Stress Scales-21. * p<.001.
|Printer friendly Cite/link Email Feedback|
|Title Annotation:||texto en ingles|
|Author:||J. Ruiz, Francisco; Suarez-Falcon, Juan Carlos; Baron-Rincon, Diego; Barrera-Acevedo, Andrea; Martin|
|Publication:||Revista Latinoamericana de Psicologia|
|Date:||Mar 22, 2016|
|Previous Article:||Alteraciones neuropsicologicas en pacientes con VIH e historia previa de consumo de sustancias. Un estudio preliminar.|
|Next Article:||Adaptacion, validacion y fiabilidad del Cuestionario de funcionamiento sexual del Hospital General de Massachusetts en una muestra colombiana y...|