Structural, Relative, and Absolute Agreement between Parents’ and Adolescent Inpatients’ Reports of Adolescent Functional Impairment

Structural, Relative, and Absolute Agreement between Parents’ and Adolescent Inpatients’ Reports of Adolescent Functional Impairment

Susan J. Frank

Susan J. Frank [1,3]

Laurie A. Van Egeren [2]

Jessica L. Fortier [1]

Paula Chase [1]

This study assessed agreement between parents’ and adolescent inpatients’ scores on caretaker and self-report versions of the Functional Impairment Scale for Children and Adolescents (FISCA and FISCA-SR). Self-report data describing impairment in eight domains were collected from 375 inpatients (M age = 15.0 years, 55% females), with parent data available for 233 (62%). Confirmatory factor analysis demonstrated structural congruence between a hypothesized, three-factor model, based on a prior study of the parent FISCA, and an observed model, based on responses to the FISCA-SR (GFJ = .95). Correlations (measuring relative agreement) and paired comparisons of means (assessing absolute agreement) generally identified stronger agreement in “public” than “private” domains of impairment, and greater relative than absolute agreement in covert, antisocial domains.

KEY WORDS: Adolescents; parents; assessment; functional impairment.

Oftentimes, parents’ and adolescents’ evaluations of adolescent problems do not concur. Some studies suggest that younger children agree more with parents than adolescents, others show the opposite, and still others observe no differences (Achenbach, McConaughy, & Howell, 1987; Edelbrock, Costello, Dulcan, Conover, & Kala, 1986; Herjanic & Reich, 1982; Moretti, Fine, Haley, & Marriage, 1985; Verhulst & van der Ende, 1992; Weissman et al., 1987); Disparate findings perhaps reflect competing developmental realities. Social cognitive advances, eventually culminating in improved perspective-taking skills, more adult-like capacities for self-reflection, and more differentiated and integrated notions of self (Harter, Bresnick, Bouchey, & Whitesell, 1997) presumably enhance the validity of adolescents’ self-evaluations (Edelbrock et al., 1986). However, gains in autonomy and more time spent alone or with peers reduce parents’ access to information regarding what or how well adolescents are doing (Brown, 1990; Larson , Richards, Moneta, Holmbeck, & Duckett, 1996). This is especially true in areas adolescents come to perceive as either “private” or “extrafamilial,” and therefore outside their parents’ jurisdiction (Smetena, 1988).

Accordingly, in this paper we argue that how agreement is defined and the domains of adolescent functioning in which it is assessed affect both the magnitude and direction of discrepancies between caretaker and adolescent informants. To demonstrate, we apply several notions of agreement to comparing parents’ and adolescent inpatients’ ratings on separate forms of the Functional Impairment Scale for Children and Adolescents (FISCA; Frank & Paul, 1995). The FISCA and its youth self-report counterpart, the FISCA-SR, assess eight domains of child problems that exemplify or contribute to inadequate social role performance. Predictions, specified according to domain, were made for three different types of agreement, with measures of each (heretofore referred to as structural, relative, and absolute) yielding different information.

In general, indices of structural agreement estimate congruence between informants’ implicit assumptions about the basic nature of the target variable. For example, goodness of fit (GFI) statistics, derived from confirmatory factor analysis (CFA), indicate the likelihood that multiple informants’ cognitive constructions of the major dimensions of the target variable converge. Alternatively, relative agreement, assessed by correlational statistics, refers to interinformant accord in ranking a target person on a specific variable relative to other members of the same population. Finally, paired comparisons of informant group means assess absolute agreement, indicated by similar estimates of the target person’s position on a given scale demarcating higher and lower levels of the target variable.

Findings from interinformant studies of child behavior problems argue for structural consistency in parents’ and adolescents’ conceptual mappings of the dimensions of adolescent psychopathology (Greenbaum, Dedrick, Prange, & Friedman, 1994; Rowe & Kandel, 1997). For example, Greenbaum et al. (1994) demonstrate that mothers,’ fathers,’ and adolescents’ scores on comparable scales of the parent and youth self-report versions of the Child Behavior Checklist (CBCL; Achenbach, 1991) contribute to the same latent variables. In contrast, studies on relative and absolute agreement suggest less concordance. Agreement coefficients hover around .25 (Achenbach et al., 1987), and evidence for sizeable differences between informant group means is commonplace (Achenbach, 1995; Young & Childs, 1994).

In recent years, research foci have shifted from assessing the magnitude of relative or absolute agreement, per se, to identifying critical moderator variables. Although an initial focus on child sex and age yielded inconsistent results, research assessing the moderating effects of the target area or domain of difficulty revealed a predictable pattern. Typically, studies of symptom and problem behavior ratings have shown that parent-adolescent agreement is weaker for self-directed, “internalizing” problems and stronger for other-directed, “externalizing” problems (Achenbach et al., 1987); a pattern that presumably could apply to ratings of functional impairment as well. However, predictions for the present study relied on a somewhat more complex model of child psychopathology.

Research Predictions

Reviewers of the child impairment literature argue increasingly for a three-dimensional model that explicitly distinguishes “overt” from “covert” forms of externalizing problems (Hinshaw, Lahey, & Hart, 1993; Tolan & Gorman-Smith, 1998), and both of these from internalizing difficulties. Recent findings offer support for a three-factor model of child functional impairment. Frank, Paul, Marks, and Van Egeren (in press) collected parent-report FISCAs for approximately 800 children and adolescents at admission to a psychiatric hospital (M age = 13.4 years, 84% 11 to 17). Using data for half their sample, they subjected scores for the eight FISCA domains to an exploratory factor analysis. CFA, applied to the remaining data, subsequently confirmed an initial, three-factor solution.

Significant pathways in Frank et al.’s (in press) observed model (shown as solid lines in Fig. 1) not only distinguished externalizing from internalizing aspects of impairment, but also overt from covert aspects of externalizing difficulties. Overt difficulties (i.e., aggression) loaded on a broader “Undercontrolled Aggression” factor; covert difficulties (i.e., delinquency and alcohol and drug use) on a “Social Role Violations” factor; and internalizing difficulties (i.e., problems related to feelings and moods) on a “Self-focused” factor. These findings were the basis for the first prediction, positing that the hypothesized structural model in Fig. 1, showing interrelationships among the eight domains of functional impairment assessed by the parent-report FISCA, adequately fits observed relationships among comparable domains assessed by adolescent inpatients’ responses to the self-report FISCA (i.e., FISCA-SR).

A priori predictions for relative and absolute agreement were made for five FISCA/FISCA-SR domains, construed as either middle or end points on a public-private continuum: Impairment stemming from aggression defined the public end; impairment revolving around inadequate regulation of internal experiences (thinking, feelings, and moods) defined the private end; and impairment associated with covert antisocial behaviors (delinquency, alcohol and drug use) was in between. Accordingly, the second prediction was that relative and absolute agreement between parents’ and adolescents’ reports are strongest for more public domains and weakest for more private domains of impairment.

The final prediction considered the direction of expected differences among informant group means. On the one hand, research suggests that parents usually report more intrapsychic difficulties for their children than children report for themselves (e.g., Kazdin, French, Unis, & Esveldt-Dawson, 1983; Verhulst & Vander Ende, 1992). Specifically, in accounting for adolescent “acting out,” parents make fewer contextual attributions than their teenage children and are more likely to attribute behavioral symptoms to a psychiatric disorder or deficit located within the adolescent (Young & Childs, 1994). On the other hand, parents appear to “underreport” problems stemming from delinquent or substance-use behaviors because adolescents typically limit their caretakers’ access to situations in which these “covert” behaviors are most likely to occur (Cantwell, Lewinsohn, Rohde, & Seeley, 1997). Accordingly, the third prediction was that adolescents evidence (a) more impaired thinking and mood difficulties when scores ar e based on parent (as compared to adolescent) report; and (b) more impairment stemming from delinquent involvements and alcohol and drug use when scores are based on adolescent (as compared to parent) report.



FISCA-SRs were collected from a total of 375 adolescent inpatients between 12 and 17 years old (M age = 15.0 years, SD = 1.4). Fifty-five percent (55%; N = 205) were females. Participants constituted 85% of all adolescents admitted over a 13-month period to a private acute psychiatric facility for children and adolescents. Although hospitalized in the same facility, this cohort of patients was admitted later than those described in Frank et al. (in press). Most (88%) were Caucasian and approximately 61% were living in two-parent (natural, adopted, or step-) families. Another 25% were in single-parent (mostly mother) homes, and the remaining 14% lived with various adult relatives or other guardians. Mean annual income, based on 330 cases and measured on an 8-point scale (0 = [less than]$8,000; 3 = $20,000 to $29,999; 4 = $30,000 to $44,999; 7 = [greater than]$70,000) was 3.6 (SD = 1.73), with approximately 6% of the households earning less than $8,000 and 13% earning $70,000 or more per year.

Caretaker data were available for 233 (62%) of the patient respondents (121 girls and 112 boys). All were admitted during 11 of the 13 months of data collection. During this time, the FISCA was part of the hospital’s intake battery, with staff collecting FISCA data from a parent or guardian (usually a mother) of 75% of all adolescent admissions. We did not use data from the other 2 study months because parent FISCAs were collected too sporadically. Respondents included 181 mothers (78%), 35 fathers (15%), 11 mother-figures (5%), and 8 father-figures (3%). Patients for whom we did (N = 233) and did not (N = 142) have parent FISCAs did not differ significantly in age, sex, ethnicity, or family income. FISCA-SR differences also were minimal.


The FISCA and FISCA-SR questionnaires (Frank & Paul, 1995) each use 172 items to assess eight domains of impairment. Items on the two forms essentially are identical in content, although referents for the target person obviously differ (i.e., “your child” vs. “I”). An objective scoring procedure links item responses to criteria identifying “Mild” (scored as 10), “Moderate” (scored as 20), or “Severe” (scored as 30) levels of impairment within each of eight domains (see the example in Table I). A child’s score within each domain is determined by the highest level at which he or she meets one or more criteria, with “No Impairment” receiving a score of 0. “Total Impairment,” ranging from 0 to 240, is computed by summing scores across domains.

Development of the FISCA criteria drew conceptually from the Child and Adolescent Functional Assessment Scale (Hodges, 1994; Hodges, Bickman, Kurtz, & Reiter, 1992). To compute internal reliability, domain criteria are treated as items, scored as 0 (not present) or 1 (present). Frank et al. (in press) reported adequate internal reliability for all but the Home scale ([alpha] = .28). Alpha coefficients for the seven other scales (each with 3 to 14 criteria) ranged from .54 to .87, with a mean of .73. Alphas for the FISCA-SR, computed for this sample, were very comparable (for Home, [alpha] = .26; for the seven other scales, .61 [less than or equal to] [alpha] [less than or equal to] .88, M = .79).

Validity data are not yet available for the FISCA-SR. However, the FISCA domains were found to discriminate inpatients from outpatients (Paul, 1997), predict hospital recidivism (Frank et al., in press), and correlate meaningfully with daily staff ratings of child and adolescent behaviors on an inpatient unit (Van Egeren, Frank, & Paul, 1999). In addition, psychological reactance (Dowd, Milne, & Wise, 1991) at hospital intake predicted changes in adolescent inpatients’ FISCA scores between discharge and 6 months follow-up (Frank et al., 1998).


Parents completed the FISCA as part of the hospital’s intake process. Adolescents completed the FISCASR within approximately 48 hours of admission in small groups supervised by one or two undergraduate externs.


Structural Agreement

We used CFA (LISREL 8.20; Joreskog & Sorbom, 1999) to assess congruence between the hypothesized and observed models. Because the chi-square likelihood ratio statistic is highly sensitive to small differences and, hence, misleading in large samples, we relied on the widely used GFI and also the Root Mean Square Error of Approximation (RMSEA) to estimate model fit. Acceptable fit was defined by a GFI [greater than] .90 and an RMSEA [less than] .08.

We ran four CFAs in all, with three applied to the FISCA-SR data for the total sample. The first tested the hypothesized model in Fig. 1 against the observed model; the second, a slightly revised model; and the third, invariance of the revised model across sex. Finally, a fourth CFA, applied to the subgroup of 233 patients with impairment data from both sources, more directly (re)assessed invariance of the original model across parent and adolescent informants. Latent factors in all models were allowed to correlate. The error variance for the Feelings and Moods Scale was set at 0 because the initial estimate was negative.

CFA Results

In general, results suggested that parents’ and adolescents’ conceptual mappings of adolescent functional impairment are structurally congruent. An initial test of the hypothesized model (N = 375) resulted in a GFI of .95 and an RMSEA of .094. Modification indices suggested a somewhat improved fit if we allowed error variances for the Delinquency and Alcohol and Drug Use scales to correlate and the Thinking Scale to load negatively on Social Role Violations. The GFI for the revised model (Fig. 2) was .98 (RMSEA = .051, .024 [less than] 90%, confidence interval [less than].077). This model proved invariant across males and females (GFI = .95; RMSEA = .050). Moreover, a final CFA, using both parent and adolescent reports (N = 233), more directly demonstrated invariance of the original hypothesized model across the two informant groups (GFI = .93; RMSEA = .075). Notably, the additional pathway from impaired thinking to Social Role Violations, emerging from the analysis performed on the self-report data alone, w as not indicated by this last analysis.

Relative and Absolute Agreement

Relative and absolute agreement were expected to be strongest for the Aggression scale and weakest for the Thinking scale and Feelings and Moods Scale, with the Delinquency and Alcohol and Drug Use scales in between. Because of the large number of analyses, statistical tests were considered significant if the probability of a type I error was less than 1%. Findings at p [less than] .05 were regarded as statistical trends.


We used regression analyses to assess whether relative or absolute agreement varied as a function of adolescent sex or age. Because none of the relevant interactions were significant, we report agreement coefficients for all 233 dyads, combined. Comparisons of agreement coefficients used a t-test for correlated data based on Downie, 1965, p. 158). [4]


t(N – 3)

= ([r.sub.a1p1] – [r.sub.a2p2]) [square root](N – 3)(1+(([r.sub.a1a2] + [r.sub.p1p2])/2))/[square root]2[1-[[r.sup.2].sub.a1p1] – [[r.sup.2].sub.a2p2] – [[r.sup.2].sub.(a1a2 + p1p2)/2] + [2r.sub.a1p1][r.sub.a2p2](([r.sub.a1a2] + [r.sub.p1p2])/2)]

where [r.sub.a1p1] equals the correlation between adolescents’ and parents’ scores for domain 1, [r.sub.a2p2] equals the correlation between adolescents’ and parents’ scores for domain 2, and ([r.sub.a1a2] + [r.sub.p1p2])/2 equals the average within-informants correlation between domain 1 and domain 2.

Although all agreement coefficients were statistically significant (p [less than] .001) they varied in magnitude. As predicted, agreement was lowest for Thinking (r(231) = .26) and Feelings and Moods (r(231) = .25). Relative agreement for each of these domains was significantly lower (p [less than] .01) than agreement for Delinquency (r(231) = .46) and Alcohol and Drug Use (r(231) = .59), and tended to be lower (p [less than] .05) than agreement for Aggression (r(231) = .43). In contrast, predicted differences between Aggression on the one hand and Delinquency and Alcohol and Drug Use on the other hand were not supported by these data. Coefficients for Aggression and Delinquency did not differ significantly, and relative agreement for Aggression tended to be weaker (p [less than] .05) rather than stronger than agreement for Alcohol and Drug Use. No predictions were made for School (r(231) = .49), Home (r(231) = .30), or Self-Harm (r(231) = .56).

Comparison of Means

Analysis of variance (ANOVA) procedures, using a 2 (Informant) x 2 (Adolescent Sex) x 2 (Adolescent Age Group; younger vs. 15 or older) design, with Informant as the within subjects factor, measured absolute levels of agreement. Results for the five domains covered by the a priori predictions are summarized in Table II. Once again, interinformant discrepancies were largest for Feelings and Moods and Thinking, followed this time by Alcohol and Drug Use, and were smallest for Aggression. Also, whereas informant differences for Delinquency were less immediately apparent, simple effects analysis of an Informant x Sex interaction (F(1,229) = 5.77, p [less than] .017) provided additional, albeit only partial, support for predicted differences in absolute agreement.

All significant parent-adolescent differences were in the direction of prediction 3. Parents’ scores exceeded adolescents’ scores for both Thinking and Feelings and Moods, whereas the reverse was true for Alcohol and Drugs. For Feelings and Moods, parent-adolescent discrepancies were significant for both sons and daughters, but larger between sons (M = 17.8, SD = 11.5) and parents (M = 26.3, SD = 8.2, t(111) = 7.4, p [less than] .001) than between daughters (M = 24.0, SD = 9.7) and parents (M = 26.8, SD = 7.2, t(120) = 28.2, p [less than] .01); for the interaction, F(1,229) = 14.32, p [less than] .001. Alternatively, parents of adolescent girls appeared to underestimate their daughters’ delinquent involvements [Ms = 14.1 (SD = 13.5) and 18.3 (SD = 13.8) respectively; t(120) = -3.19, p [less than] .01], whereas this was not the case for parents of adolescent boys (Ms for both sources were around 22.5; for the Informant x Sex interaction, F(1,229) = 5.77, p [less than] .017).


In general, these data supported the predictions for structural and absolute agreement and partially supported the prediction for relative agreement. In the earlier study with the parent measure, identification of a three-factor model, anticipated from the extant literature(e.g., Hinshaw et al., 1993; Tolan & Gorman-Smith, 1998), was cited as evidence for the FISCA’s construct validity (Frank et al., in press). By extension, structural congruence between the latent model identified by Frank et al. and the observed model identified in the FISCA-SR data advocates for the construct validity of the self-report FISCA as well. Furthermore, the one notable distinction between the hypothesized and observed models, found for Thinking, adds rather than detracts from this line of reasoning.

Whereas we correctly predicted that impaired thinking would load positively on the Undercontrolled Aggression factor, we did not anticipate it loading in the opposite direction on the Social Role Violations factor. These two findings can be viewed as logical counterparts: Several different theories of child psychopathology cohere to the hypothesis that poorly differentiated and confused thinking contributes to undercontrolled aggression (Frank & Quinlan, 1976; Hinshaw et al., 1993; Lochman & Dodge, 1994; Lyons-Ruth, 1996). Furthermore, its converse–that instrumental as compared to undercontrolled forms of antisocial behavior demand more effective reality testing–has been suggested as well (e.g., Frank & Quinlan, 1976; Lyons-Ruth, 1996).

Although it is not yet clear why only the first assumption (and not its converse) emerged in analyses involving the parent data, overall, isomorphism in parents’ and adolescents’ conceptual mappings of adolescent functional impairment justified subsequent tests of relative and absolute agreement. One caveat in interpreting this second set of results: caretaker data were unavailable for 38% of the patient respondents, raising the issue of unknown sampling biases linked to unmeasured variables (no such biases were evident for variables we did assess).

Interestingly, relative agreement was stronger and more consistent than anticipated, and the exceptions were predicted: that is, agreement was more mediocre when parents and adolescents reported on the adolescents’ difficulties regulating their private thoughts and feelings. In contrast, greater discrepancy in informants’ relative rankings of covert as opposed to overt dimensions of antisocial problems did not emerge, and if anything, relative agreement in rating impairment associated with covert forms of deviance tended to exceed consensus in ratings of overt aggression.

Alternatively, broader support for the a priori predictions was evident when agreement was defined in “absolute” terms. First, interinformant discrepancies were largest for domains of impairment associated with the adolescents’ subjective experiences, with the direction of disagreement consistent with the hypothesis that parents, more than adolescents, prefer to “locate” adolescent problems “within” the adolescent. Second, whereas parents and adolescents similarly estimated the adolescents’ level of aggression, adolescents acknowledged more dysfunction stemming from their use of alcohol and drugs, and adolescent girls reported more dysfunction in the form of delinquent activities.

Underreporting of delinquency by parents of girls, not found for parents of boys, may reflect the influence of sex-role stereotypes, leading parents to deny certain sex-atypical problems in their children (Tarullo et al., 1995). However, the alternative hypothesis–that parents are less tolerant of and hence more likely to report child psychopathology that violates gender-linked social role expectations (Jensen, Traylor, Xenakis, & Davis, 1988; Tarullo et al., 1995)–is more consistent with the findings for feelings and moods. For both sexes, parents attributed more emotional impairment to adolescents than adolescents attributed to themselves; however, these discrepancies were significantly greater when parents were describing their sons’ as compared to their daughters’ mood difficulties.

Juxtaposing specific findings for relative and absolute agreement within a particular domain illustrates ways in which alternative definitions can lead to different appraisals of cross-informant agreement. Consider, for example, the relatively large difference in means, alongside a relatively strong correlation between parent and adolescent scores on the Alcohol and Drug Use Scale. Perhaps cohort effects, along with parents’ obliviousness to these sorts of covert activities, lead parents to underestimate their teenagers’ actual level of substance use-related difficulties. Nevertheless, disparities in absolute agreement do not preclude quite satisfactory levels of relative agreement. In spite of intergenerational disparities in reported levels of difficulty, parents appear sensitive to how their adolescent children’s covert antisocial involvements measure up to those of their peers (Whitbeck, Hoyt, & Ackley, 1997).

Achenbach (1995) touched on the clinical implications of findings like these. He argues that requiring the same absolute threshold for all informants in determining a child’s “caseness” creates unnecessary obstacles to agreement. However, rules for assigning different cutoffs to different groups of informants must be founded in well-replicated studies that take into account the area of difficulty being assessed.

In a different vein, the adolescents’ gender and age appear to be less important as moderators of parent- adolescent agreement than either type of agreement or domain of impairment. More generally, we suspect that other variables contributed to the unexpectedly strong consensus between parents’ and adolescents’ rankings of adolescent problems. The fact that items and criteria associated with many of the FISCA domains are anchored in concrete events (e.g., expulsion from school, a police contact or arrest, losing a job because of drinking or drug use) or specific behaviors (e.g., a lethal suicide attempt, kicking or hitting a peer) may partially explain why correlations for more than half the FISCA domains well exceeded the magnitude of association typically reported in this literature (Achenbach et al., 1987). However, the use of an inpatient sample may have contributed as well. In particular, adolescent difficulties that are sufficiently chronic or severe to require hospitalization are likely to be more obv ious to parents than adolescent difficulties more toward the mild end of the impairment continuum. Although this possibility could not be investigated here, Achenbach et al.’s (1987) review suggests somewhat better interinformant agreement in seriously disturbed samples of children or adolescents than in normal or outpatient populations.

Whether the findings will replicate across diverse facilities as well as at various levels of care should be addressed directly, as should the relative utility of parent as opposed to adolescent report in predicting outcomes of adolescents’ functional difficulties. These concerns not withstanding, this study succeeded in illustrating the value of considering alternative notions of agreement and varying the domain of functioning in which agreement is assessed. A greater demand for multiple informant studies in part results from, and at the same time, raises questions about discrepant reporting by adolescent and adult informants. In making sense of an oftentimes confusing literature, it is important to distinguish among structural, relative, and absolute notions of agreement, and also between private and public aspects of adolescents’ experiences, and those aspects that adolescents and adults typically encounter together and apart.


The authors want to thank Rivedell of Michigan Hospital, as well as the many adolescents and parents who made this research possible.

(1.) Department of Psychology, Michigan State University, East Lansing, MI.

(2.) Department of Psychology, Institute of Children, Youth, and Families, Michigan State University, East Lansing, MI.

(3.) Corresponding author, Department of Psychology, Michigan State University, East Lansing, MI, 48824. Phone: 517-355-1832. Fax: 517-432-2476. e-mail:


Achenbach, T. M. (1991). Integrative guide for the 1991 CBCL/4-18, YSR, and TRF profiles. Burlington, University of Vermont, Department of Psychiatry.

Achenbach, T. M. (1995). Diagnosis, assessment, and comorbidity in psychosocial treatment research. Journal of Abnormal Child Psychology, 23, 45-65.

Achenbach, T. M., MeConaughy, S. H., & Howell, C. T. (1987). Child/adolescent behavioral and emotional problems. Implications of cross-informant correlations for situational specificity. Psychological Bulletin, 101, 2 13-232.

Brown, B. (1990). Peer groups. In S. Feldman & G. Elliott (Eds.), At the threshold: The developing adolescent (pp. 17 1-196). Cambridge, MA: Harvard University Press.

Cantwell, D. P., Lewinsohn, M. D., Rohde, P., & Seeley, J. R. (1997). Correspondence between adolescent report and parent report of psychiatric diagnostic data. Journal of the American Academy of Child and Adolescent Psychiatry, 36, 610-619.

Dowd, E., Milne, C., & Wise, S. (1991). The Therapeutic Reactance Scale: A measure of psychological reactance. Journal of Counseling and Development, 69, 541-545.

Edelbrock, C., Costello, A. J., Dulcan, M. K., Conover, N. C., & Kala, R. (1986). Parent-child agreement on child psychiatric symptoms assessed via structured interview. Journal of Child Psychology and Psychiatry, 27, 181-190.

Frank, S. J., & Paul, J. S. (1995). The Functional Impairment Scale for Children and Adolescents. Department of Psychology, Michigan State University.

Frank, S. J., Jackson-Walker, S., Marks, M., Van Egeren, L. A., Loop, K., & Olson, K. (1998). From the laboratory to the hospital, adults to adolescents, and disorders to personality: The case of psychological reactance. Journal of Clinical Psychology 54, 361-381.

Frank, S. J., Paul, J. S., Marks, M., & Van Egeren, L. A. (in press). Initial validation of the functional impairment scale for children and adolescents. Journal of the American Academy of Child and Adolescent Psychiatry.

Frank, S. J., & Quinlan, D. M. (1976). Ego development and female delinquency: A cognitive-developmental approach. Journal of Abnormal Psychology, 85, 505-510.

Greenbaum, P. E., Dedrick, R. F., Prange, M. E., & Friedman, R. M. (1994). Parent, teacher, and child ratings of problem behaviors of youngsters with serious emotional disturbances. Psychological Assessment, 6, 141-148.

Harter, S., Bresnick, S., Bouchey, H. A., & Whitesell, N. R. (1997). The development of multiple role-related selves during adolescence. Development and Psychopathology, 9, 835-853.

Herjanic, B., & Reich, W. (1982). Development of a structured psychiatric interview for children. Agreement between child and parent on individual symptoms. Journal of Abnormal Child Psychology, 10, 307-324.

Hinshaw, S. P., Lahey, B. B., & Hart, E. L. (1993). Issues of taxonomy and comorbidity in the development of conduct disorder. Development and Psychopathology, 5, 31-49.

Hodges, K. (1994). The Child and Adolescent Functional Assessment Scale. Unpublished measure, Department of Psychology, Eastem Michigan University, Ypsilanti, MI.

Hodges, K., Bickman, L., Kurtz, S., & Reiter, M. (1992). A multidimensional measure of level of functioning for children and adolescents. In A. Algarin & R. M. Friedman (Eds.), A system of care for children’s mental health: Building a research hose, Florida Mental Health Institute, University of South Florida.

Jensen, P. S., Traylor, J., Xenakis, S. N., & Davis, H. (1988). Child psychopathology rating scales and interrater agreement: I. Parents’ gender and psychiatric symptoms. Journal of the American Academy of Child and Adolescent Psychiatry, 27, 442-450.

Joreskog, K. G., & Sorbom, D. (1999). Lisrel 8.20. Chicago: Scientific Software International, Inc.

Kazdin, A. E., French, N. H., Unis, A. S., & Esveldt-Dawson, K. (1983). Assessment of childhood depression: Correspondence of child and parent ratings. Journal of the American Academy of Child Psychiatry, 22, 157-164.

Larson, R. W., Richards, M. H., Moneta, G., Holmbeck, G., & Duckett, E. (1996). Changes in adolescents’ daily interactions with their families from ages 10 to 18: Disengagement and transformation. Developmental Psychology 32, 744-754.

Lochman, J. E., & Dodge, K. A. (1994). Social-cognitive processes of severely violent, moderately aggressive, and non-aggressive boys. Journal of Consulting and Clinical Psychology 62, 366-374.

Lyons-Ruth, K. (1996). Attachment relationships among children with aggressive behavior problems: The role of disorganized early attachment patterns. Journal of Consulting and Clinical Psychology, 64, 64-73.

Moretti, M. M., Fine, S., Haley, G., & Marriage, K. (1985). Childhood and adolescent depression. Child-report versus parent-report information. Journal of the American Academy of Child Psychiatry 24, 298-302.

Paul, J. A. (1997). Validational study of the Functional Impairment Scale for Children and Adolescents, Presented at the Biennial Meetings of the Society for Research on Child Development, Washington DC.

Rowe, D. C., & Kandel, D. (1997). In the eye of the beholder? Parental ratings of externalizing and internalizing symptoms. Journal of Abnormal Child Psychology 25, 263-275.

Smetena, J. (1988). Concepts of self and social convention: Adolescents’ and parents’ reasoning about hypothetical and actual family conflicts. In M. Gunnar (Ed.), 21st Minnesota symposium on child psychology (pp. 79-122). Hillsdale, NJ: Erlbaum.

Tarullo, L. B., Richardson, D. T., Radke-Yarrow, M., & Martinez, P. E. (1995). Multiple sources in child diagnosis: Parent-child concordance in affectively ill and well families. Journal of Clinical Child Psychology, 24, 173-183.

Tolan, P. H., & Gorman-Smith, D. (1998). Development of serious and violent offender careers. In R. Loeber & D. P. Farrington (Eds.), Serious and violent juvenile offenders: Risk factors and successful interventions, Ch. 5, pp. 68-85. Thousand Oaks, CA.: Sage.

Van Egeren, L. A., Frank, S. J., & Paul, 3. 5. (1999). Daily behavior ratings among child and adolescent inpatients: The Abbreviated Child Behavior Rating Form (CBRF-A). Journal of the American Academy of Child and Adolescent Psychiatry, 38, 1412-1425.

Verhulst, F. C., & van der Ende, J. (1992). Agreement between parents’ reports and adolescents’ self-reports of problem behavior. Journal of Child Psychology and Psychiatry 56, 909-915.

Weissman, M. M., Wickramaratne, P., Warner, V. John, K., Prusoff, B. A., Merikangas, K. R., & Gammon, D. (1987). Assessing psychiatric disorders in children. Archives of General Psychiatry, 44, 747-753.

Whitbeck, L. B., Hoyt, D. R., & Ackley, K. A. (1997). Families of homeless and runaway adolescents: A comparison of parent/caretaker and adolescent perspectives on parenting, family violence, and adolescent conduct. Child Abuse and Neglect, 21, 517-528.

Young, D. W., & Childs, N. A. (1994). Family images of hospitalized adolescents: The failure to generate shared understandings. Psychiatry 57, 258-257.

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