In their review of the literature, Roberts et al. (1998)
concluded that prevalence rates from studies using earlier versions of the DISC and DSM criteria were in the 18–20% range. Our overall prevalence rate is clearly in this range. There actually appears to be even more congruence between recent studies and ours. Canino et al. (2004)
note that when their sample was restricted to those 9–17, and similar diagnoses included for Puerto Rico and the Great Smoky Mountains Study, the overall prevalence in Puerto Rico was 17.3% and the rate in the latter study (Angold et al., 2002
) was 17.7%. Again, our prevalence for 11–17 year-olds was 17.1%. Other studies have reported higher rates and also lower rates. McGee and colleagues (McGee et al., 1990
; Feehan et al., 1994
) have reported prevalences of DSM-III disorder of 22% for 15-year-olds in New Zealand and 36% for DSM-III-R disorders when the cohort was 18. On the other hand, Lewinsohn et al. (1993)
reported a point prevalence of 11% for DSM-III-R disorders among 14–18 year-olds, while Ford et al. (2003)
reported a prevalence of 9.5%. In the MECA study, Shaffer et al. (1996)
reported a prevalence based on youth report of 32.2%, 15.3% with CGAS score of ≤ 70 and 12.3% with DISC and CGAS impairment.
Recently, Simpson et al. (2005)
report that approximately 5% of youths 4–17 years of age in the U.S. had emotional or behavioral problems (by parent report) and about 80% of these were functionally impaired. Thus, about 4% had symptoms and impairment. Their research did not use structured diagnostic interviews nor DSM diagnostic criteria nor were youths interviewed.
We should note that our 1-year prevalence of agoraphobia makes this the most prevalent anxiety disorder. No other papers have reported rates of agoraphobia using the DSM-IV with adolescents. In their review, Black et al. (2004)
do not report data on agoraphobia, nor does the chapter specifically address issues of reliability or validity of this diagnosis. The chapter does cite rates of panic disorder and social phobia similar to what we report. DSM-IV-TR (American Psychiatric Association) notes that it is difficult to diagnose agoraphobia due to frequent overlap with specific phobias, and notes that agoraphobia without panic is more prevalent than panic disorder with agoraphobia. This is what we found as well. It appears that the DISC-IV may be overly sensitive with regards to agoraphobia, but absent data from other studies, we can only report what we found. Our overall rate of anxiety disorder (6.9%) was lower than the 9.5% reported by Canino et al. (2004)
, who excluded agoraphobia. Their highest rate of anxiety disorder was separation anxiety (5.7%), but their sample was age 4–17. This issue clearly awaits additional data from other studies.
Similar to the findings from other studies of children and adolescents, without adjusting for impairment, we found many of the expected associations of psychiatric disorders with gender and age (Costello et al., 1997
; Ford et al., 2003
; Feehan et al., 1994
; Lewinsohn et al., 1993
; Bird et al., 1989
; Simpson et al., 2005
) without adjusting for impairment. Similar to other studies, we did not find strong associations of disorders with indicators of socioeconomic status (see Costello et al., 1997
; Canino et al., 2004
Our results adjusting for impairment yielded similar effects to those noted by others (Costello et al., 1996
; Shaffer et al., 1996
; Bor et al., 1997
; Simonoff et al., 1997
; Narrow et al., 1998
). In our study, adjusting for DISC-IV impairment reduced prevalence 54%. Adjusting for CGAS score reduced prevalence threefold (to 5.4). Based on their review, Roberts et al. (1998)
concluded that prevalence of disorder was probably 7–12%, adjusting for impairment. Our rates were 5.3 to 11%, depending on the impairment criteria used. These are highly consistent with rates adjusted for impairment reported by Costello et al. (2003)
, Canino et al. (2004)
, and Ford et al. (2003)
. In a reanalysis of existing datasets, Costello et al. (1998)
found that the median prevalence of disorder adjusted for global impairment was 5.4% (the same as our rate with CGAS ≤ 69) with a range of 4.3% to 7.4%.
Adjusting for impairment eliminated initial effects of family income and caregiver education (already minimal). The effect of CGAS score ≤ 69 was greater than for the DISC algorithm for moderate impairment. However, adjusting for neither DISC impairment nor CGAS score appreciably altered the observed associations between disorders and age, gender, or marital status of caregiver. Overall, the most consistent effects were observed for gender, age, and marital status of the primary caregiver, with or without impairment. Our results, which need to be replicated, suggest that the epidemiology of disorders may vary, depending on the degree of impairment.
However, there remains no consensus on how to operationalize clinically important impairment in epidemiologic studies (Roberts et al., 1998
; Costello et al., 1997
; Simonoff et al., 1997
; Narrow et al., 1998
). No doubt lack of such consensus contributes to discordant results across studies. Although work is proceeding in this regard (Angold et al., 1999
), much remains to be done, particularly in the case of children and adolescents. The CGAS is an example of an assessment strategy that has proven useful. Its utility to date is limited due to the fact that it has not been widely used in epidemiologic studies and there is variation in its application across studies.
Questions might arise from our sample design. We did not select an area probability sample. In an attempt to compensate for this design effect, we post-stratified the TH2K sample to approximate the age, gender and ethnic distribution of the 5-county metropolitan area in which all of our study households were located. We used the DISC-IV and DSM-IV diagnostic criteria, with and without adjustment for impairment. We would argue that the high concordance between our prevalence rates and other studies discussed above provide strong evidence for the external validity of our results. We might add that numerous studies cited by Roberts et al. (1998)
and studies by Shaffer et al. (1996)
and Turner and Gil (2002)
were not, in a study of 19–21 year olds, strictly speaking, area probability samples either.
Another issue related to our sampling strategy involves the socioeconomic composition of the sample. Our sample was underrepresented by families below poverty levels extant in the metropolitan area where subjects resided. This may explain, in part, the fact that we did not find much evidence for an association between youth psychiatric disorders and family income or education. However, the relation between SES and mental health has proven complex and not consistent across studies of adults (see Holzer et al., 1986
; Dohrenwend, 1990
; Dohrenwend et al., 1992
). To illustrate this with data from adolescents, Johnson et al. (1999)
found evidence for adverse effects of parental SES on a number of adolescent psychiatric disorders, prospectively. The sample was mostly European American. Wadsworth and Achenbach (2005)
also found lower SES increased risk of a range of mental health problems (but not all). Their sample was mostly majority youths and no data were presented separately for minority youths. We also have found evidence for increased odds of youth disorders when only data from European Americans are examined (Roberts et al., 2006
), but not for African or Mexican American youths. This result for minority youths has been reported from other studies. Costello et al. (1997)
found poverty increased the odds of disorders among majority but not minority youths. Canino et al. (2004)
found no association between family income and youth disorders in Puerto Rico. Thus, evidence for the role of family SES remains unclear in the risk of child and adolescent psychiatric disorders.
A second issue is that we did not interview parents about the DSM-IV disorders assessed by youth interview. As noted by Roberts et al. (1998)
, a substantial proportion of studies have relied on either parent or youth report, but not both. In the Oregon Adolescent Depression Study, Lewinsohn et al. (1993)
relied only on adolescent self-reports as did Turner and Gil (2002)
more recently for older adolescents. And while there is argument that data from multiple informants is desirable, many studies have demonstrated considerable discordance in parent-child reports of psychopathology, whether subjects are assessed with structured (Edelbrock et al., 1986
; Jensen et al., 1999
) or semi-structured interviews (Angold et al., 1987
), using symptom dimensions or categorical diagnoses (Rubio-Stipec et al., 1994
), or using checklists (Achenbach et al., 1987
; Yeh et al., 2001
). There is little consensus on how parent and youth reports should be combined in epidemiologic studies and, indeed, some authors report prevalences separately for parents and children.
In conclusion, the evidence we have presented suggests that the prevalence findings for children and adolescents are remarkably robust across diverse setting, using diverse methods to generate estimates of DSM-IV caseness. At this point, data from U.S. rural samples, the island of Puerto Rico, a sample from England, and our sample from a large, metropolitan area of the U.S. yield similar estimates of prevalences, both with and without adjustment for impairment. Our next step is to examine the natural history of DSM-IV disorders in the TH2K and risk factors for disorders over time using data from 2 waves of observation 12 months apart. This will provide much-needed prospective data on the epidemiology of psychiatric disorders among adolescents in the United States.