J. psychiat. Res., Vol. 25, No. 3, pp. 141-151, 1991. Printed in Great Britain.
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TRENDS ARTIFACT
MEREDITH G .
WARSHAW,t'2
IN MAJOR
DEPRESSION
AN
OF RECALL?
GERALD L . KLERMAN 3 a n d PHILIP W . LAVORI 1
I Department of Psychiatry and H u m a n Behavior, Brown University, Providence, RI 02906, U.S.A and 3Department of Psychiatry, Cornell University Medical College, New York, U.S.A. (Received 13 August 1990; revised 9 January 1991) Summary--There is evidence that rates of major depression have increased over this century, with successive birth cohorts showing increased lifetime risks and earlier ages of onset. Two memory effects have been considered possible artifactual causes of these trends: age-related forgetting and postdating early episodes. In this study, relatives were reinterviewed six years after study entry using interviewers blind to initial reports. We examined the stability of lifetime diagnoses of MDD and ages of first onset. Older relatives were no more likely than younger ones to lose diagnoses nor to postdate their ages of first MDD onset. This is evidence that memory artifacts are not solely responsible for the observed secular trends.
Introduction THERE is growing evidence that the rates of major depression have been increasing over this century, with successive birth cohorts showing increased lifetime risks and earlier ages of onset (Gershon, Hamovit, Guroff, & Nurnberger, 1987; Klerman et al., 1985; Weissman, Kidd, & Prosuff, 1982; Bland & Orn, 1986; Wittchen, 1986; Hallstrom, 1984; Robins et al., 1984; Weissman & Myers, 1978; Hagnell, 1982). Discussions of these secular trends have reviewed possible artifacts that could explain the observed secular trends in lifetime risk and age of onset (Klerman et al., 1985; Lavori et al., 1987; Klerman & Weissman, 1989; Hasin & Link, 1988). Two kinds of memory effects have been considered especially likely: (1) age-related forgetting of earlier episodes, and (2) remembering first onsets of affective disorders as being more recent than they really are (postdating). Age-related forgetting of earlier episodes could lead to an apparent secular trend in rates of major depression (MDD) if older relatives who had MDD at an early age were more likely to forget those episodes than were younger ones. This artifact could also lead to an apparent change in age of onset over successive cohorts, with younger cohorts showing earlier ages of onset, if older relatives with multiple episodes were more likely to forget their early episodes. Postdating could lead to the apparent period effect by causing ages of first onset to be remembered as being close to the interview date, whether or not they really were. Lavori et al. (1987) eliminated the effects of memory loss associated with advanced age by limiting their sample to relatives born after 1929, and still found the secular trend. 2Address for correspondence: Meredith G. Warshaw, Butler Hospital, Duncan Building, 345 Blackstone Blvd., Providence, RI 02906, U.S.A. 141
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Previous work (Rice, Endicott, Knesevich, & Rochberg, 1987) found that stability of diagnosis was predicted by receiving medication, number of symptoms, and number of episodes. Using the number of symptoms to create an index of "caseness", Rice (1989) examined secular trends using different levels of caseness, varying severity. He found that although the secular trends were diminished with increasing casesness, they remained statistically significant. Similar findings have been reported by Endicott (1984). Other possible artifacts such as the sample selection effects of differential mortality and institutionalization were discussed in Lavori et al. (1987). These selection artifacts exert their hypothetical effects by reducing apparent rates in older relatives. However, the locus of the observed secular trend is not in low rates in the elderly, but in the surprisingly high rates in younger cohorts. In ~he N I M H - C R B Collaborative Depression Study (CDS) we reinterviewed relatives six years after their initial entry into the study, using interviewers blind to initial reports. Having two independent assessments six years apart permitted us to examine the amounts of changing diagnoses and ages of first onset of MDD, and to examine whether changes occurred most frequently in older relatives. In this paper, we examine the concordance of diagnoses between the initial interview (T1) and the blind reassessment six years later (T2), as well as the agreement between reported ages of first onset for relatives diagnosed with major depressive disorder (MDD) at both interviews. We then look at whether these changes can account for the observed secular trends. Methods The N I M H - C R B Collaborative Depression Study is a long-term prospective follow-up study of patients who sought treatment for major affective disorders at one of five collaborating medical centers in Boston, Chicago, Iowa City, New York City, and St. Louis (Katz & Klerman, 1979). Their spouses and first-degree relatives (parents, siblings and children) aged 17 or older were asked to participate in a family study, and those who agreed were interviewed with the S A D S - L and other standardized instruments (Spitzer et al., 1978; Katz & Klerman, 1979). Interviews were conducted during the years 1978-82 (T1). Diagnoses of m a j o r depressive disorder were made using the Research Diagnostic Criteria (RDC) (Spitzer, Endicott, & Robins, 1978). Six years after the initial interviews (T2), relatives were again assessed by interviewers blind to the original reports at T1. This paper focuses on the lifetime diagnoses of M D D in 1,684 relatives inteviewed at both T1 and T2 (1,001 women and 683 men), after considering the effects of missing data at T2.
Assessing Agreement There are several ways for someone's assessment to change from T l to T2: the person's lifetime diagnosis can be (1) positive at T1 and dropped at T2, (2) positive at T1 and positive at T2 with a change in onset date that is still pre-T1, (3) positive at T1 and positive at T2 but with a post-T1 onset date, (4) negative at T1 and positive at T2 with a pre-T1 onset date, or (5) negative at T1 and positive at T2 with a post-T1 onset date. The first four of these cases are clearly inconsistent with each other, while the last case may be either correct or an undetected error. Without a gold standard for either diagnosis or date of
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onset, we cannot tell whether a dropped diagnosis means a failure to detect a true episode or the correction o f a false positive diagnosis. Symmetrically, we cannot tell whether a T2 positive diagnosis that is inconsistent with the T1 diagnosis is correct or incorrect. Therefore, we attempt to infer the characteristics of the diagnostic measurement process by studying the outcomes of repeated measurements: stability of diagnosis and onset date. In this paper we examine both changes in diagnosis and onset date, each of which present special statistical challenges. Kappas were calculated to assess the agreement of T1 and T2 ratings of M D D onset before T1. For any relative, an M D D diagnosis at T2 could have an onset date either before or after T1 (see Table 1). For the purpose of assessing T 1 - T 2 agreement on the question of "Did this person have M D D by TI?", a relative who was not diagnosed as having M D D at the T1 assessment and was given an M D D diagnosis at T2 with an onset date after T1 was considered a new onset, and therefore an agreement (both interviewers agreed that
Table 1 MDD Diagnoses at TI and T2 A. All relatives MDD MDD TI
No No 911 Yes 143 Total 1054 K = . 57
(79%) (27%)
T2 Yes (before TI) 103 (9%) 341 (64%) 444 observed = .82 expected = .59
Yes (after T1) 134 (12%) 52 (10%) 186
Total 1148 536
Yes (after T1) 89 (14%) 36 (10%) 125
Total 630 371
Yes (after T1) 45 (9%) 16 (10%) 61
Total 518 165
B. Women MDD MDD T1
No No 485 Yes 92 Total 577 K = .59
(77%) (25%)
T2 Yes (before T1) 56 (9%) 243 (66%) 299 observed = .82 expected = .55
C. Men MDD MDD T1
No No 426 Yes 51 Total 477 K = .52
(82%) (31%)
T2 Yes (before T1) 47 (9%) 98 (59%) 145 observed = .83 expected = .65
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M.G. WARSHAWet al.
MDD did not occur before TI). A relative who was given a diagnosis of MDD at TI and at T2, but whose T2 date of onset was after T1, was considered a disagreement (since the T2 interviewer had concluded that this person had not ever had MDD before T1). To estimate agreement on date of first onset of MDD, we only considered those relatives with positive diagnoses at both T1 and T2. In other words, our analyses of changes of onset date are conditional on being diagnosed with MDD on two occasions. The intraclass correlation coefficient (Snedecor & Cochran, 1967) is often used to summarize agreement on continuous measures such as age of onset. We have used it to assess agreement in onset dates from T1 to T2, although we recognize that the peculiarities of this situation (including the conditioning) may recommend other ways of looking at stability. As in the case of the kappa statistic discussed above, no single number summary is completely adequate in this situation. Since reports at T2 have six more years of possible range of answers than at T1, the temporal asymmetry and difference in truncation complicate interpretation of the intra-class correlation coefficient, as does the conditioning on repeated diagnosis. For these reasons, we also examined graphs of changes in reported ages of onset to better understand the complexities of the stability of onset dates. Results
Missing Data Relatives could be lost to this interviewed sample at two points: the original T1 assessment or the T2 blind reassessment. At T1, all relatives had a diagnosis made by F H - R D C by two informants and approximately half were also directly interviewed using the S A D S - L . Lavori, Keller, and Endicott (1988) examined the rates of MDD in relatives not interviewed at T1, as assessed by family history interviews (FH-RDC). They proposed using explicit modelling and imputation in conjunction with the F H - R D C results in order to avoid the problem of underdiagnosis when relying solely on family informant data. In this model, they used information about various strata in the interviewed sample to model the likely rates in the uninterviewed group. They found evidence that the rates of MDD in uninterviewed relatives were strongly associated with age, sex, and relationship to proband, and that stratification by these variables could help to reduce bias due to differential likelihood of interview. Their results suggest that, as sex and cohort predict both diagnosis and likelihood of interview, stratifying on those variables (as is done throughout the following analyses) reduces the amount of confounding. Because the F H - R D C s were not repeated at T2, they do not give us information on diagnostic stability; therefore in the remainder of this paper we only consider the relatives interviewed at T1. Five hundred and fifty-two (25°70) of the relatives interviewed at T1 were not interviewed at T 2 : 1 1 % due to refusal to participate a second time, 5% had died in the time since T1, probands refused permission for 3% to be interviewed, 2% could not be located, 2% could not be interviewed (mostly because of organic problems such as advanced senility), and 2% were not interviewed for other miscellaneous reasons. Women were more likely than men to be reinterviewed (79% versus 71%, ;~2= 18.06, d f = 1, p < .0001). In addition, younger cohorts were more likely than older to be reinterviewed (Xz= 43.60, d f = 5, p < .0001), with rates of not being reinterviewed ranging
SECULAR TRENDS IN MAJOR DEPRESSION
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from 54o'/0 in those born before 1906 to 18% in the cohort born in 1956-65. There was no significant difference in rates of lifetime diagnosis of MDD at T1 between relatives not interviewed at T2 and those reinterviewed, within strata defined by sex and birth cohort, execept in the two youngest cohorts of women where the reinterviewed relatives had a lower rate of MDD. Because those not interviewed at T2 give no information on stability, we concentrate in this paper on those relatives assessed at both T1 and T2.
Transitions in Diagnostic Status from T1 to T2 Using the decision rule discussed in the Methods section above, agreement was good (kappa = .57) for the sample as a whole (see Table 1). The women's diagnoses are slightly more stable than the men's, with kappas of .59 and .52 respectively. In Table 1, "observed" refers to the proportion of subjects for whom the diagnoses received at T1 and T2 agree. " E x p e c t e d " refers to the proportion for whom agreement would be expected by chance alone, given the observed marginal rates. Of the 536 relatives who were given lifetime diagnoses of MDD at T1, 143 (27°70) were not given those diagnoses at T2 (see Table 1). The rates of dropped diagnoses did not differ significantly across strata defined by sex or cohort (x 2 (19)= 18.33, p = .50). Similarly, the rates of T1 MDD diagnoses receiving post-T1 onset dates at T2 did not differ significantly by sex/cohort strata (X2 (19) = 12.24, p = .88), occurring in 10°70 of the TI MDD diagnoses for both men and women. For the relatives with no lifetime diagnosis of MDD at T1, 103 (9°70) received diagnoses of MDD at T2 with onset dates before T1. This rate was identical for men and women, and did not differ significantly by cohort (X2= 26.34, d f = 19, p = . 12). It cannot be known for sure how many of these were really new onsets in the period T1 to T2 with an age of onset given at the T2 interview that was backdated, as opposed to disagreement over whether there was actually an episode of MDD during the period covered during the T1 assessment ( " b a c k d a t i n g " versus "spontaneous generation" of episodes).
Changes in Marginal Rates of Diagnostic Status from T1 to T2 In contrast to the similarity of the transition probabilities for men and women across cohorts, the graphs of "disease-free" survival probabilities by age for men and women, stratified by cohort, show that women tend to have a net decrease in numbers of diagnoses of MDD at T2 (see Figure 1). The age-specific "survival" disease-free probabilities are actually higher at T2 than T1 for two thirds of the women (all but the two youngest cohorts, those born 1946-65), indicating that the net number of diagnoses is decreasing. By contrast, for the men, there is no consistent pattern of difference except for the cohort born in 1926-35, which had a net decrease in diagnoses. The lack of consistent pattern found for the men is consistent with errors randomly distributed over cohort and interview occasion. The rates of diagnoses "disappearing" at T2 or showing up " n e w " at T2 with pre-T1 onset dates were examined in order to try to understand the patterns in the survival curves. For the sample as a whole, 143 (27°7o) of the 536 cases of MDD diagnosed at T1 had "disappeared" at T2 while 103 (21°70) of the 444 MDD cases diagnosed at T2 with pre-T1 dates of onset (T2/pre-T1) were not diagnosed at T1, i.e. were " n e w " at T2 (Table la). For the men, these two figures were almost identical, with 51 (31°70) of the 165 T1 MDD
146
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SECULAR TRENDS IN MAJOR DEPRESSION
147
cases disappearing and 47 (32°7o) o f the 145 T2 cases of M D D with pre-T1 onset dates (T2/pre-T1) being new (Table lb). For women, however, there was a large discrepancy in these figures, with 92 (25O7o) of the 371 T1 M D D disappearing but only 56 (19°70) of the 299 T2/pre-T1 M D D being new (Table lc). In other words, for men the number of apparent onsets before T1 that disappeared at T2 and the number that spontaneously appeared at that time were about the same, but for women only about 65°7o as m a n y discordant cases appeared as disappeared at T2.
Comparison of Marginal and Transitional Points of View Men and women both had T1 M D D transition probabilities for diagnoses disappearing at rates that are not statistically different and also both had about the same odds of having new T2/pre-T 1 diagnoses appear. Yet, for the women, the ratio of new T2/pre-T1 diagnoses appearing to diagnoses disappearing is approximately 1:1.6 while for the men it is 1:1.1. This occurs because the women have a higher marginal rate of T1 M D D to begin with (37°70 vs. 24°70, p < .0001). Because the proportion of women having M D D at T1 was higher than the proportion of men, the same odds of not repeating a T1 diagnosis lead to a larger proportion of the total sample of women losing their diagnoses. Similarly, since a smaller proportion of the women were diagnosed as MD-free to begin with, they ended up gaining a smaller proportion of new T2/pre-T1 diagnoses. The combination of these two factors led to a larger proportion of the women's diagnoses disappearing. This explains the difference between the survival graphs (which show marginal rates) for the men and women.
Changes in Diagnosis by Period o f First Onset The question arises as to which diagnoses are being dropped. If the problem is age-related forgetting, we would expect to see the episodes with the earliest onset dates being lost at higher rates than the more recent ones. We therefore divided the sample by period of onset into approximately equal size groups and examined the rates of dropped diagnoses by period (see Table 2). Thirty-eight percent of M D D episodes with onset dates in the most recent period, 1976-81, were dropped as opposed to an average of 23°7o for the other periods (z = 3.0, p < .003). Further, this was not a gradual drop, decreasing over successive periods, but rather was a sharp step, with the rate remaining approximately constant once the episode was more than one period back from T1, i.e. before 1976. Table 2
Rates of Dropped Diagnoses by Period Period 1976-81 1970-75 1958-69 1928-57
No. of cases MDD at TI
No. o f cases dropped at T2
% Dropped
140 163 155 78
53 37 33 20
38°7o 23 °7o 21% 26%
Onset Dates The level of agreement o f ages of first onsets (for those receiving M D D diagnoses at
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M.G. WARSHAWet al.
both T1 and T2) f r o m initial interview to reassessment was g o o d . For the 363 relatives reporting lifetime episodes o f M D D at both T1 and T2, the intra-class correlation for age o f first onset was .82. The intra-class correlation o f onset ages given at T1 and T2 for relatives receiving lifetime diagnoses o f M D D at both assessments are very similar for m e n and w o m e n (.79 versus .83, respectively). The m e a n change in age o f onset f r o m T1 to T2 was 1.4 year later ( m o r e recent) at T2. H o w e v e r , the m e d i a n change was 0, with 50% o f the relatives changing their ages o f onset between - 1 and 3 years. Figure 2 is a scatterplot
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o f the ages o f first onset o f M D D f r o m T1 versus ages f r o m T2. A line is drawn through T1 = T2. In addition, a lowess (robust locally weighted regression) (Cleveland, 1979) line is drawn through the scatterplot and, except at the very oldest and youngest ages, is quite close to the line T1 = T2.
149
SECULAR TRENDS IN MAJOR DEPRESSION
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There is no evidence of the changes in age of onset being related to age at interview; in particular, older relatives did not change their ages of onset more than younger relatives, nor were their changes in a consistently different direction. In Figure 3, the relatives are grouped into four age categories by quartiles, and box plots drawn for the change in age from T1 to T2. In these plots, a horizontal line is drawn through the box at the median of the data and the ends of the box are at the upper and lower quartiles. The vertical lines go up and down from the box to the points 1.5 times the interquartile range away from the quartiles; points further than that are drawn individually. The distributions of the changes in age, in particular medians and interquartile ranges, are very similar for the four groups. In other words, there is no evidence that older relatives were postdating their onset ages more than younger ones. Discussion The large secular trends in MDD onset found in our dataset and others provoke two kinds of q u e s t i o n s - - " I s the effect real or artifactual?" and " W h a t can it m e a n ? " . In this paper we have tried to answer a form of the first question: " C a n the simplest artifacts,
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M.G. WARSHAWet al.
f o r g e t t i n g a n d p o s t d a t i n g , e x p l a i n the o b s e r v e d secular t r e n d ? " O u r d a t a suggest t h a t the a n s w e r is n o t s t r a i g h t f o r w a r d ; a l t h o u g h these a r t i f a c t s m a y have s o m e role, t h e y d o not tell the w h o l e s t o r y , as t h e y d o n o t seem to be solely r e s p o n s i b l e for the secular t r e n d s in the d a t a . O n t h e o t h e r h a n d , o u r results p o i n t o u t s o m e l i m i t a t i o n s o f the lifetime diagnosis method of studying major depression. O u r results d o n o t s u p p o r t the i d e a t h a t the a p p a r e n t secular t r e n d is solely the result o f d i f f e r e n t i a l f o r g e t t i n g b y o l d e r c o h o r t s . W e f o u n d t h a t o l d e r relatives were no m o r e likely to lose diagnoses in the interval f r o m T1 to T2 t h a n were the y o u n g e r ones. Therefore, a l t h o u g h we d i d find evidence o f d i a g n o s t i c instability, it does n o t seem to be r e s p o n s i b l e for the a g e - p e r i o d - c o h o r t effect. In a d d i t i o n , the m o s t recent episodes were the ones m o s t likely to be d r o p p e d , r a t h e r t h a n the oldest. W e c a n n o t tell f r o m these d a t a w h e t h e r this is b e c a u s e is it m o r e difficult to d i f f e r e n t i a t e a true M D D e p i s o d e f r o m a less severe d e p r e s s i o n when s o m e o n e is in it o r close to it, or whether any e p i s o d e m e m o r a b l e e n o u g h to be r e m e m b e r e d for a few years is m e m o r a b l e e n o u g h to be r e m e m b e r e d f o r a lifetime. O n e p o s s i b l e i n t e r p r e t a t i o n is t h a t each event has its o w n p r o b a b i l i t y o f being d i a g n o s e d as a depression, with that p r o b a b i l i t y d e c r e a s i n g shortly after the event a n d t h e n leveling o f f . These results i n d i c a t e t h a t lifetime d i a g n o s i s is u n s t a b l e as a c a t e g o r i c a l v a r i a b l e . In c o n t r a s t , given a g r e e m e n t o n the existence o f M D D , the t i m i n g o f first onset s h o w e d g o o d reliability f r o m o r i g i n a l interview to b l i n d reassessment. F u r t h e r , there was no evidence o f a n y m a j o r systematic d i r e c t i o n o f c h a n g e in the ages o f onset r e p o r t e d at T2 versus T1. In p a r t i c u l a r , o l d e r relatives were no m o r e likely t h a n y o u n g e r ones to p o s t d a t e their ages o f first M D D onset. This is evidence t h a t the a r t i f a c t o f " p o s t d a t i n g " is p r o b a b l y n o t solely r e s p o n s i b l e for the o b s e r v e d secular trends. W e realize that six years is a short interval for answering questions a b o u t lifetime changes in m e m o r y . W i t h a longer interval, such as 2 0 - 3 0 years, it is likely that the rates o f agreement would worsen a n d possible that the predictors o f stability might change. This is an i m p o r t a n t q u e s t i o n to a d d r e s s a n d a c o m p e l l i n g r e a s o n for studies with longer f o l l o w - u p . W e need to f u r t h e r e x p l o r e w a y s assessing causes o f d i a g n o s t i c instability. F o r instance, the relatives w h o s e M D D d i a g n o s e s were stable over the t w o interviews m i g h t represent the m o r e severe cases, o r they might just be the people who are the most reliable i n f o r m a n t s . This i n f o r m a t i o n will p r o v e i m p o r t a n t for p s y c h i a t r i c genetics, where the a c c u r a c y o f each individual diagnosis is crucial. Rice et al. (1987) have f o u n d that diagnostic stability is related to receiving m e d i c a t i o n , n u m b e r o f s y m p t o m s , a n d n u m b e r o f episodes. It will be i m p o r t a n t to f u r t h e r s t u d y the influences o n stability b o t h in o r d e r to i m p r o v e it a n d to e x a m i n e its effects o n results in p s y c h i a t r i c research, as well as to successfully p u r s u e the a g e n d a s o f p s y c h i a t r i c e p i d e m i o l o g y a n d genetics.
Acknowledgements--From the National Institute of Mental Health Collaborative Program on the Psychobiology of Depression--Clinical Studies. conducted with the participation of the following investigators: G. L. Klerman (Chairperson) (New York); R. M. A. Hirschfeld (Co-Chairperson) (Washington, DC); M. B. Keller and P. Lavori (Boston); J. A. Fawcett and W. A. Scheftner (Chicago); W. CoryeU, N. C. Andreasen, J. Haley and P. Wasek (Iowa City); J. Endicott and J. E. Loth (New York); J. Rice and T. Reich (St. Louis). Other contributors include: P. J. Clayton, J. Courghan, M. M. Katz, E. Robins, R. W. Shapiro, R. L. Spitzer and G. Winokur.
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References Bland, R., & Orn, H. (1986). Family violence and psychiatric disorder. Canadian Journal o f Psychiatry 31, 129-137. Cleveland, W. (1979). Robust locally weighted regression and smoothing scatterplots. Journal o f the American Statistical Association 74, 829-826. Endicott, J. (1984). Sex differences in symptom reporting. Presented at Analysis of Cohort Effect in Family and Population Data, Boston, MA. Fleiss, J. (1981). Statistical methods f o r rates and proportions. New York: John Wiley & Sons. Gershon, E., Hamovit, J., Guroff, J., & Nurnberger, J. (1987). Birth cohort changes in manic and depressive disorders in relatives of bipolar and schizoaffective affective patients. Archives o f General Psychiatry 44, 314-319. Hagnell, O. (1982). The 25 year follow-up of the Lundby Study: Incidence and risk of alcoholism, depression and disorders of the senium. In J. Barrett & R. Rose (Eds.), Mental disorders in the community. New York, Guilford Press. Hallstrom, T. (1984). Point prevalence of major depressive disorder in a Swedish urban female population. Acta Psychiatrica Scandinavica 69, 52-69. Hasin, D., & Link, B. (1988). Age and recognition of depression: Implications for a cohort effect in major depression. Psychological Medicine 18, 683-688. Kalbfleisch, J., & Prentice, R. (1980). The Statistical analysis o f failure time data. New York: John Wiley & Sons. Katz, M., & Klerman, G. (1979). Introduction: Overview of the clinical studies program. American Journal o f Psychiatry 136, 49-51. Klerman, G., Lavori, P., Rice, J., Reich, T., Endicott, J., Andreasen, N., Keller, M., & Hirschfield, R. (1985). Birth-cohort trends in rates of major depressive disorder among relatives of patients with affective disorder. Archives o f General Psychiatry 42, 689-693. Klerman, G., & Weissman, M. (1989). Increasing rates of depression. Journal o f the American MedicalAssociation 261, 2229-2235. Lavori, P., Keller, M., & Endicott, J. (1988). Improving the validity of FH-RDC diagnosis of major affective disorder in uninterviewed relatives in family studies: A model based approach. Journal o f Psychiatric Research 22, 249-259. Lavori, P., Klerman, G., Keller, M., Reich, T., Rice, J., & Endicott, J. (1987). Age-period-cohort analysis of secular trends in onset of major depression: Findings in siblings of patients with major affective disorder. Journal o f Psychiatric Research 21, 23-35. Rice, J. (1989). Stability of diagnoses in relatives. Depression among non-patients: A 6year follow-up. Symposium conducted at the 142nd Annual Meeting of the American Psychiatric Association, San Francisco. Rice, J., Endicott, J., Knesevich, M., & Rochberg, N. (1987). The estimation of diagnostic sensitivity using stability data: An application to major depressive disorder. Journal o f Psychiatric Research 21, 337-345. Robins, L., Helzer, J., Weissman, M., Orvashel, H., Gruenberg, E., Burke, J., & Regier, D. (1984). Lifetime prevalence of specific psychiatric disorders in three sites. Archives o f General Psychiatry 41, 949-958. Snedecor, G., & Cochran, W. (1967). Statistical methods. Iowa: The Iowa State University Press. Spitzer, R., Endicott, J., & Robins, E. (1978). Research diagnostic criteria (3rd ed.). New York: Biometrics Research, New York State Department of Mental Hygiene. Weissman, M., Kidd, K., & Prosuff, B. (1982). Variability in rates of affective disorder in relatives of depressed and normal probands. Archives o f General Psychiatry 29, 1397-1403. Weissman, M., & Myers, J. (1978). Affective disorders in a U.S. urban community: The use of Research Diagnostic Criteria in a community survey. Archives o f General Psychiatry 35, 1304-1311. Wittchen, H. (1986). Contribution of epidemiological data to the classification of anxiety disorders. In I. Hand & H. Wittchen (Eds.), Panic phobias. Berlin: Springer.
ARE
SECULAR
0022-3956/91 $3.00 + .00 Pergamon Press plc
TRENDS ARTIFACT
MEREDITH G .
WARSHAW,t'2
IN MAJOR
DEPRESSION
AN
OF RECALL?
GERALD L . KLERMAN 3 a n d PHILIP W . LAVORI 1
I Department of Psychiatry and H u m a n Behavior, Brown University, Providence, RI 02906, U.S.A and 3Department of Psychiatry, Cornell University Medical College, New York, U.S.A. (Received 13 August 1990; revised 9 January 1991) Summary--There is evidence that rates of major depression have increased over this century, with successive birth cohorts showing increased lifetime risks and earlier ages of onset. Two memory effects have been considered possible artifactual causes of these trends: age-related forgetting and postdating early episodes. In this study, relatives were reinterviewed six years after study entry using interviewers blind to initial reports. We examined the stability of lifetime diagnoses of MDD and ages of first onset. Older relatives were no more likely than younger ones to lose diagnoses nor to postdate their ages of first MDD onset. This is evidence that memory artifacts are not solely responsible for the observed secular trends.
Introduction THERE is growing evidence that the rates of major depression have been increasing over this century, with successive birth cohorts showing increased lifetime risks and earlier ages of onset (Gershon, Hamovit, Guroff, & Nurnberger, 1987; Klerman et al., 1985; Weissman, Kidd, & Prosuff, 1982; Bland & Orn, 1986; Wittchen, 1986; Hallstrom, 1984; Robins et al., 1984; Weissman & Myers, 1978; Hagnell, 1982). Discussions of these secular trends have reviewed possible artifacts that could explain the observed secular trends in lifetime risk and age of onset (Klerman et al., 1985; Lavori et al., 1987; Klerman & Weissman, 1989; Hasin & Link, 1988). Two kinds of memory effects have been considered especially likely: (1) age-related forgetting of earlier episodes, and (2) remembering first onsets of affective disorders as being more recent than they really are (postdating). Age-related forgetting of earlier episodes could lead to an apparent secular trend in rates of major depression (MDD) if older relatives who had MDD at an early age were more likely to forget those episodes than were younger ones. This artifact could also lead to an apparent change in age of onset over successive cohorts, with younger cohorts showing earlier ages of onset, if older relatives with multiple episodes were more likely to forget their early episodes. Postdating could lead to the apparent period effect by causing ages of first onset to be remembered as being close to the interview date, whether or not they really were. Lavori et al. (1987) eliminated the effects of memory loss associated with advanced age by limiting their sample to relatives born after 1929, and still found the secular trend. 2Address for correspondence: Meredith G. Warshaw, Butler Hospital, Duncan Building, 345 Blackstone Blvd., Providence, RI 02906, U.S.A. 141
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Previous work (Rice, Endicott, Knesevich, & Rochberg, 1987) found that stability of diagnosis was predicted by receiving medication, number of symptoms, and number of episodes. Using the number of symptoms to create an index of "caseness", Rice (1989) examined secular trends using different levels of caseness, varying severity. He found that although the secular trends were diminished with increasing casesness, they remained statistically significant. Similar findings have been reported by Endicott (1984). Other possible artifacts such as the sample selection effects of differential mortality and institutionalization were discussed in Lavori et al. (1987). These selection artifacts exert their hypothetical effects by reducing apparent rates in older relatives. However, the locus of the observed secular trend is not in low rates in the elderly, but in the surprisingly high rates in younger cohorts. In ~he N I M H - C R B Collaborative Depression Study (CDS) we reinterviewed relatives six years after their initial entry into the study, using interviewers blind to initial reports. Having two independent assessments six years apart permitted us to examine the amounts of changing diagnoses and ages of first onset of MDD, and to examine whether changes occurred most frequently in older relatives. In this paper, we examine the concordance of diagnoses between the initial interview (T1) and the blind reassessment six years later (T2), as well as the agreement between reported ages of first onset for relatives diagnosed with major depressive disorder (MDD) at both interviews. We then look at whether these changes can account for the observed secular trends. Methods The N I M H - C R B Collaborative Depression Study is a long-term prospective follow-up study of patients who sought treatment for major affective disorders at one of five collaborating medical centers in Boston, Chicago, Iowa City, New York City, and St. Louis (Katz & Klerman, 1979). Their spouses and first-degree relatives (parents, siblings and children) aged 17 or older were asked to participate in a family study, and those who agreed were interviewed with the S A D S - L and other standardized instruments (Spitzer et al., 1978; Katz & Klerman, 1979). Interviews were conducted during the years 1978-82 (T1). Diagnoses of m a j o r depressive disorder were made using the Research Diagnostic Criteria (RDC) (Spitzer, Endicott, & Robins, 1978). Six years after the initial interviews (T2), relatives were again assessed by interviewers blind to the original reports at T1. This paper focuses on the lifetime diagnoses of M D D in 1,684 relatives inteviewed at both T1 and T2 (1,001 women and 683 men), after considering the effects of missing data at T2.
Assessing Agreement There are several ways for someone's assessment to change from T l to T2: the person's lifetime diagnosis can be (1) positive at T1 and dropped at T2, (2) positive at T1 and positive at T2 with a change in onset date that is still pre-T1, (3) positive at T1 and positive at T2 but with a post-T1 onset date, (4) negative at T1 and positive at T2 with a pre-T1 onset date, or (5) negative at T1 and positive at T2 with a post-T1 onset date. The first four of these cases are clearly inconsistent with each other, while the last case may be either correct or an undetected error. Without a gold standard for either diagnosis or date of
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onset, we cannot tell whether a dropped diagnosis means a failure to detect a true episode or the correction o f a false positive diagnosis. Symmetrically, we cannot tell whether a T2 positive diagnosis that is inconsistent with the T1 diagnosis is correct or incorrect. Therefore, we attempt to infer the characteristics of the diagnostic measurement process by studying the outcomes of repeated measurements: stability of diagnosis and onset date. In this paper we examine both changes in diagnosis and onset date, each of which present special statistical challenges. Kappas were calculated to assess the agreement of T1 and T2 ratings of M D D onset before T1. For any relative, an M D D diagnosis at T2 could have an onset date either before or after T1 (see Table 1). For the purpose of assessing T 1 - T 2 agreement on the question of "Did this person have M D D by TI?", a relative who was not diagnosed as having M D D at the T1 assessment and was given an M D D diagnosis at T2 with an onset date after T1 was considered a new onset, and therefore an agreement (both interviewers agreed that
Table 1 MDD Diagnoses at TI and T2 A. All relatives MDD MDD TI
No No 911 Yes 143 Total 1054 K = . 57
(79%) (27%)
T2 Yes (before TI) 103 (9%) 341 (64%) 444 observed = .82 expected = .59
Yes (after T1) 134 (12%) 52 (10%) 186
Total 1148 536
Yes (after T1) 89 (14%) 36 (10%) 125
Total 630 371
Yes (after T1) 45 (9%) 16 (10%) 61
Total 518 165
B. Women MDD MDD T1
No No 485 Yes 92 Total 577 K = .59
(77%) (25%)
T2 Yes (before T1) 56 (9%) 243 (66%) 299 observed = .82 expected = .55
C. Men MDD MDD T1
No No 426 Yes 51 Total 477 K = .52
(82%) (31%)
T2 Yes (before T1) 47 (9%) 98 (59%) 145 observed = .83 expected = .65
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M.G. WARSHAWet al.
MDD did not occur before TI). A relative who was given a diagnosis of MDD at TI and at T2, but whose T2 date of onset was after T1, was considered a disagreement (since the T2 interviewer had concluded that this person had not ever had MDD before T1). To estimate agreement on date of first onset of MDD, we only considered those relatives with positive diagnoses at both T1 and T2. In other words, our analyses of changes of onset date are conditional on being diagnosed with MDD on two occasions. The intraclass correlation coefficient (Snedecor & Cochran, 1967) is often used to summarize agreement on continuous measures such as age of onset. We have used it to assess agreement in onset dates from T1 to T2, although we recognize that the peculiarities of this situation (including the conditioning) may recommend other ways of looking at stability. As in the case of the kappa statistic discussed above, no single number summary is completely adequate in this situation. Since reports at T2 have six more years of possible range of answers than at T1, the temporal asymmetry and difference in truncation complicate interpretation of the intra-class correlation coefficient, as does the conditioning on repeated diagnosis. For these reasons, we also examined graphs of changes in reported ages of onset to better understand the complexities of the stability of onset dates. Results
Missing Data Relatives could be lost to this interviewed sample at two points: the original T1 assessment or the T2 blind reassessment. At T1, all relatives had a diagnosis made by F H - R D C by two informants and approximately half were also directly interviewed using the S A D S - L . Lavori, Keller, and Endicott (1988) examined the rates of MDD in relatives not interviewed at T1, as assessed by family history interviews (FH-RDC). They proposed using explicit modelling and imputation in conjunction with the F H - R D C results in order to avoid the problem of underdiagnosis when relying solely on family informant data. In this model, they used information about various strata in the interviewed sample to model the likely rates in the uninterviewed group. They found evidence that the rates of MDD in uninterviewed relatives were strongly associated with age, sex, and relationship to proband, and that stratification by these variables could help to reduce bias due to differential likelihood of interview. Their results suggest that, as sex and cohort predict both diagnosis and likelihood of interview, stratifying on those variables (as is done throughout the following analyses) reduces the amount of confounding. Because the F H - R D C s were not repeated at T2, they do not give us information on diagnostic stability; therefore in the remainder of this paper we only consider the relatives interviewed at T1. Five hundred and fifty-two (25°70) of the relatives interviewed at T1 were not interviewed at T 2 : 1 1 % due to refusal to participate a second time, 5% had died in the time since T1, probands refused permission for 3% to be interviewed, 2% could not be located, 2% could not be interviewed (mostly because of organic problems such as advanced senility), and 2% were not interviewed for other miscellaneous reasons. Women were more likely than men to be reinterviewed (79% versus 71%, ;~2= 18.06, d f = 1, p < .0001). In addition, younger cohorts were more likely than older to be reinterviewed (Xz= 43.60, d f = 5, p < .0001), with rates of not being reinterviewed ranging
SECULAR TRENDS IN MAJOR DEPRESSION
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from 54o'/0 in those born before 1906 to 18% in the cohort born in 1956-65. There was no significant difference in rates of lifetime diagnosis of MDD at T1 between relatives not interviewed at T2 and those reinterviewed, within strata defined by sex and birth cohort, execept in the two youngest cohorts of women where the reinterviewed relatives had a lower rate of MDD. Because those not interviewed at T2 give no information on stability, we concentrate in this paper on those relatives assessed at both T1 and T2.
Transitions in Diagnostic Status from T1 to T2 Using the decision rule discussed in the Methods section above, agreement was good (kappa = .57) for the sample as a whole (see Table 1). The women's diagnoses are slightly more stable than the men's, with kappas of .59 and .52 respectively. In Table 1, "observed" refers to the proportion of subjects for whom the diagnoses received at T1 and T2 agree. " E x p e c t e d " refers to the proportion for whom agreement would be expected by chance alone, given the observed marginal rates. Of the 536 relatives who were given lifetime diagnoses of MDD at T1, 143 (27°70) were not given those diagnoses at T2 (see Table 1). The rates of dropped diagnoses did not differ significantly across strata defined by sex or cohort (x 2 (19)= 18.33, p = .50). Similarly, the rates of T1 MDD diagnoses receiving post-T1 onset dates at T2 did not differ significantly by sex/cohort strata (X2 (19) = 12.24, p = .88), occurring in 10°70 of the TI MDD diagnoses for both men and women. For the relatives with no lifetime diagnosis of MDD at T1, 103 (9°70) received diagnoses of MDD at T2 with onset dates before T1. This rate was identical for men and women, and did not differ significantly by cohort (X2= 26.34, d f = 19, p = . 12). It cannot be known for sure how many of these were really new onsets in the period T1 to T2 with an age of onset given at the T2 interview that was backdated, as opposed to disagreement over whether there was actually an episode of MDD during the period covered during the T1 assessment ( " b a c k d a t i n g " versus "spontaneous generation" of episodes).
Changes in Marginal Rates of Diagnostic Status from T1 to T2 In contrast to the similarity of the transition probabilities for men and women across cohorts, the graphs of "disease-free" survival probabilities by age for men and women, stratified by cohort, show that women tend to have a net decrease in numbers of diagnoses of MDD at T2 (see Figure 1). The age-specific "survival" disease-free probabilities are actually higher at T2 than T1 for two thirds of the women (all but the two youngest cohorts, those born 1946-65), indicating that the net number of diagnoses is decreasing. By contrast, for the men, there is no consistent pattern of difference except for the cohort born in 1926-35, which had a net decrease in diagnoses. The lack of consistent pattern found for the men is consistent with errors randomly distributed over cohort and interview occasion. The rates of diagnoses "disappearing" at T2 or showing up " n e w " at T2 with pre-T1 onset dates were examined in order to try to understand the patterns in the survival curves. For the sample as a whole, 143 (27°7o) of the 536 cases of MDD diagnosed at T1 had "disappeared" at T2 while 103 (21°70) of the 444 MDD cases diagnosed at T2 with pre-T1 dates of onset (T2/pre-T1) were not diagnosed at T1, i.e. were " n e w " at T2 (Table la). For the men, these two figures were almost identical, with 51 (31°70) of the 165 T1 MDD
146
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SECULAR TRENDS IN MAJOR DEPRESSION
147
cases disappearing and 47 (32°7o) o f the 145 T2 cases of M D D with pre-T1 onset dates (T2/pre-T1) being new (Table lb). For women, however, there was a large discrepancy in these figures, with 92 (25O7o) of the 371 T1 M D D disappearing but only 56 (19°70) of the 299 T2/pre-T1 M D D being new (Table lc). In other words, for men the number of apparent onsets before T1 that disappeared at T2 and the number that spontaneously appeared at that time were about the same, but for women only about 65°7o as m a n y discordant cases appeared as disappeared at T2.
Comparison of Marginal and Transitional Points of View Men and women both had T1 M D D transition probabilities for diagnoses disappearing at rates that are not statistically different and also both had about the same odds of having new T2/pre-T 1 diagnoses appear. Yet, for the women, the ratio of new T2/pre-T1 diagnoses appearing to diagnoses disappearing is approximately 1:1.6 while for the men it is 1:1.1. This occurs because the women have a higher marginal rate of T1 M D D to begin with (37°70 vs. 24°70, p < .0001). Because the proportion of women having M D D at T1 was higher than the proportion of men, the same odds of not repeating a T1 diagnosis lead to a larger proportion of the total sample of women losing their diagnoses. Similarly, since a smaller proportion of the women were diagnosed as MD-free to begin with, they ended up gaining a smaller proportion of new T2/pre-T1 diagnoses. The combination of these two factors led to a larger proportion of the women's diagnoses disappearing. This explains the difference between the survival graphs (which show marginal rates) for the men and women.
Changes in Diagnosis by Period o f First Onset The question arises as to which diagnoses are being dropped. If the problem is age-related forgetting, we would expect to see the episodes with the earliest onset dates being lost at higher rates than the more recent ones. We therefore divided the sample by period of onset into approximately equal size groups and examined the rates of dropped diagnoses by period (see Table 2). Thirty-eight percent of M D D episodes with onset dates in the most recent period, 1976-81, were dropped as opposed to an average of 23°7o for the other periods (z = 3.0, p < .003). Further, this was not a gradual drop, decreasing over successive periods, but rather was a sharp step, with the rate remaining approximately constant once the episode was more than one period back from T1, i.e. before 1976. Table 2
Rates of Dropped Diagnoses by Period Period 1976-81 1970-75 1958-69 1928-57
No. of cases MDD at TI
No. o f cases dropped at T2
% Dropped
140 163 155 78
53 37 33 20
38°7o 23 °7o 21% 26%
Onset Dates The level of agreement o f ages of first onsets (for those receiving M D D diagnoses at
148
M.G. WARSHAWet al.
both T1 and T2) f r o m initial interview to reassessment was g o o d . For the 363 relatives reporting lifetime episodes o f M D D at both T1 and T2, the intra-class correlation for age o f first onset was .82. The intra-class correlation o f onset ages given at T1 and T2 for relatives receiving lifetime diagnoses o f M D D at both assessments are very similar for m e n and w o m e n (.79 versus .83, respectively). The m e a n change in age o f onset f r o m T1 to T2 was 1.4 year later ( m o r e recent) at T2. H o w e v e r , the m e d i a n change was 0, with 50% o f the relatives changing their ages o f onset between - 1 and 3 years. Figure 2 is a scatterplot
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o f the ages o f first onset o f M D D f r o m T1 versus ages f r o m T2. A line is drawn through T1 = T2. In addition, a lowess (robust locally weighted regression) (Cleveland, 1979) line is drawn through the scatterplot and, except at the very oldest and youngest ages, is quite close to the line T1 = T2.
149
SECULAR TRENDS IN MAJOR DEPRESSION
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There is no evidence of the changes in age of onset being related to age at interview; in particular, older relatives did not change their ages of onset more than younger relatives, nor were their changes in a consistently different direction. In Figure 3, the relatives are grouped into four age categories by quartiles, and box plots drawn for the change in age from T1 to T2. In these plots, a horizontal line is drawn through the box at the median of the data and the ends of the box are at the upper and lower quartiles. The vertical lines go up and down from the box to the points 1.5 times the interquartile range away from the quartiles; points further than that are drawn individually. The distributions of the changes in age, in particular medians and interquartile ranges, are very similar for the four groups. In other words, there is no evidence that older relatives were postdating their onset ages more than younger ones. Discussion The large secular trends in MDD onset found in our dataset and others provoke two kinds of q u e s t i o n s - - " I s the effect real or artifactual?" and " W h a t can it m e a n ? " . In this paper we have tried to answer a form of the first question: " C a n the simplest artifacts,
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M.G. WARSHAWet al.
f o r g e t t i n g a n d p o s t d a t i n g , e x p l a i n the o b s e r v e d secular t r e n d ? " O u r d a t a suggest t h a t the a n s w e r is n o t s t r a i g h t f o r w a r d ; a l t h o u g h these a r t i f a c t s m a y have s o m e role, t h e y d o not tell the w h o l e s t o r y , as t h e y d o n o t seem to be solely r e s p o n s i b l e for the secular t r e n d s in the d a t a . O n t h e o t h e r h a n d , o u r results p o i n t o u t s o m e l i m i t a t i o n s o f the lifetime diagnosis method of studying major depression. O u r results d o n o t s u p p o r t the i d e a t h a t the a p p a r e n t secular t r e n d is solely the result o f d i f f e r e n t i a l f o r g e t t i n g b y o l d e r c o h o r t s . W e f o u n d t h a t o l d e r relatives were no m o r e likely to lose diagnoses in the interval f r o m T1 to T2 t h a n were the y o u n g e r ones. Therefore, a l t h o u g h we d i d find evidence o f d i a g n o s t i c instability, it does n o t seem to be r e s p o n s i b l e for the a g e - p e r i o d - c o h o r t effect. In a d d i t i o n , the m o s t recent episodes were the ones m o s t likely to be d r o p p e d , r a t h e r t h a n the oldest. W e c a n n o t tell f r o m these d a t a w h e t h e r this is b e c a u s e is it m o r e difficult to d i f f e r e n t i a t e a true M D D e p i s o d e f r o m a less severe d e p r e s s i o n when s o m e o n e is in it o r close to it, or whether any e p i s o d e m e m o r a b l e e n o u g h to be r e m e m b e r e d for a few years is m e m o r a b l e e n o u g h to be r e m e m b e r e d f o r a lifetime. O n e p o s s i b l e i n t e r p r e t a t i o n is t h a t each event has its o w n p r o b a b i l i t y o f being d i a g n o s e d as a depression, with that p r o b a b i l i t y d e c r e a s i n g shortly after the event a n d t h e n leveling o f f . These results i n d i c a t e t h a t lifetime d i a g n o s i s is u n s t a b l e as a c a t e g o r i c a l v a r i a b l e . In c o n t r a s t , given a g r e e m e n t o n the existence o f M D D , the t i m i n g o f first onset s h o w e d g o o d reliability f r o m o r i g i n a l interview to b l i n d reassessment. F u r t h e r , there was no evidence o f a n y m a j o r systematic d i r e c t i o n o f c h a n g e in the ages o f onset r e p o r t e d at T2 versus T1. In p a r t i c u l a r , o l d e r relatives were no m o r e likely t h a n y o u n g e r ones to p o s t d a t e their ages o f first M D D onset. This is evidence t h a t the a r t i f a c t o f " p o s t d a t i n g " is p r o b a b l y n o t solely r e s p o n s i b l e for the o b s e r v e d secular trends. W e realize that six years is a short interval for answering questions a b o u t lifetime changes in m e m o r y . W i t h a longer interval, such as 2 0 - 3 0 years, it is likely that the rates o f agreement would worsen a n d possible that the predictors o f stability might change. This is an i m p o r t a n t q u e s t i o n to a d d r e s s a n d a c o m p e l l i n g r e a s o n for studies with longer f o l l o w - u p . W e need to f u r t h e r e x p l o r e w a y s assessing causes o f d i a g n o s t i c instability. F o r instance, the relatives w h o s e M D D d i a g n o s e s were stable over the t w o interviews m i g h t represent the m o r e severe cases, o r they might just be the people who are the most reliable i n f o r m a n t s . This i n f o r m a t i o n will p r o v e i m p o r t a n t for p s y c h i a t r i c genetics, where the a c c u r a c y o f each individual diagnosis is crucial. Rice et al. (1987) have f o u n d that diagnostic stability is related to receiving m e d i c a t i o n , n u m b e r o f s y m p t o m s , a n d n u m b e r o f episodes. It will be i m p o r t a n t to f u r t h e r s t u d y the influences o n stability b o t h in o r d e r to i m p r o v e it a n d to e x a m i n e its effects o n results in p s y c h i a t r i c research, as well as to successfully p u r s u e the a g e n d a s o f p s y c h i a t r i c e p i d e m i o l o g y a n d genetics.
Acknowledgements--From the National Institute of Mental Health Collaborative Program on the Psychobiology of Depression--Clinical Studies. conducted with the participation of the following investigators: G. L. Klerman (Chairperson) (New York); R. M. A. Hirschfeld (Co-Chairperson) (Washington, DC); M. B. Keller and P. Lavori (Boston); J. A. Fawcett and W. A. Scheftner (Chicago); W. CoryeU, N. C. Andreasen, J. Haley and P. Wasek (Iowa City); J. Endicott and J. E. Loth (New York); J. Rice and T. Reich (St. Louis). Other contributors include: P. J. Clayton, J. Courghan, M. M. Katz, E. Robins, R. W. Shapiro, R. L. Spitzer and G. Winokur.
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References Bland, R., & Orn, H. (1986). Family violence and psychiatric disorder. Canadian Journal o f Psychiatry 31, 129-137. Cleveland, W. (1979). Robust locally weighted regression and smoothing scatterplots. Journal o f the American Statistical Association 74, 829-826. Endicott, J. (1984). Sex differences in symptom reporting. Presented at Analysis of Cohort Effect in Family and Population Data, Boston, MA. Fleiss, J. (1981). Statistical methods f o r rates and proportions. New York: John Wiley & Sons. Gershon, E., Hamovit, J., Guroff, J., & Nurnberger, J. (1987). Birth cohort changes in manic and depressive disorders in relatives of bipolar and schizoaffective affective patients. Archives o f General Psychiatry 44, 314-319. Hagnell, O. (1982). The 25 year follow-up of the Lundby Study: Incidence and risk of alcoholism, depression and disorders of the senium. In J. Barrett & R. Rose (Eds.), Mental disorders in the community. New York, Guilford Press. Hallstrom, T. (1984). Point prevalence of major depressive disorder in a Swedish urban female population. Acta Psychiatrica Scandinavica 69, 52-69. Hasin, D., & Link, B. (1988). Age and recognition of depression: Implications for a cohort effect in major depression. Psychological Medicine 18, 683-688. Kalbfleisch, J., & Prentice, R. (1980). The Statistical analysis o f failure time data. New York: John Wiley & Sons. Katz, M., & Klerman, G. (1979). Introduction: Overview of the clinical studies program. American Journal o f Psychiatry 136, 49-51. Klerman, G., Lavori, P., Rice, J., Reich, T., Endicott, J., Andreasen, N., Keller, M., & Hirschfield, R. (1985). Birth-cohort trends in rates of major depressive disorder among relatives of patients with affective disorder. Archives o f General Psychiatry 42, 689-693. Klerman, G., & Weissman, M. (1989). Increasing rates of depression. Journal o f the American MedicalAssociation 261, 2229-2235. Lavori, P., Keller, M., & Endicott, J. (1988). Improving the validity of FH-RDC diagnosis of major affective disorder in uninterviewed relatives in family studies: A model based approach. Journal o f Psychiatric Research 22, 249-259. Lavori, P., Klerman, G., Keller, M., Reich, T., Rice, J., & Endicott, J. (1987). Age-period-cohort analysis of secular trends in onset of major depression: Findings in siblings of patients with major affective disorder. Journal o f Psychiatric Research 21, 23-35. Rice, J. (1989). Stability of diagnoses in relatives. Depression among non-patients: A 6year follow-up. Symposium conducted at the 142nd Annual Meeting of the American Psychiatric Association, San Francisco. Rice, J., Endicott, J., Knesevich, M., & Rochberg, N. (1987). The estimation of diagnostic sensitivity using stability data: An application to major depressive disorder. Journal o f Psychiatric Research 21, 337-345. Robins, L., Helzer, J., Weissman, M., Orvashel, H., Gruenberg, E., Burke, J., & Regier, D. (1984). Lifetime prevalence of specific psychiatric disorders in three sites. Archives o f General Psychiatry 41, 949-958. Snedecor, G., & Cochran, W. (1967). Statistical methods. Iowa: The Iowa State University Press. Spitzer, R., Endicott, J., & Robins, E. (1978). Research diagnostic criteria (3rd ed.). New York: Biometrics Research, New York State Department of Mental Hygiene. Weissman, M., Kidd, K., & Prosuff, B. (1982). Variability in rates of affective disorder in relatives of depressed and normal probands. Archives o f General Psychiatry 29, 1397-1403. Weissman, M., & Myers, J. (1978). Affective disorders in a U.S. urban community: The use of Research Diagnostic Criteria in a community survey. Archives o f General Psychiatry 35, 1304-1311. Wittchen, H. (1986). Contribution of epidemiological data to the classification of anxiety disorders. In I. Hand & H. Wittchen (Eds.), Panic phobias. Berlin: Springer.
