A Longitudinal Study of Suicidal Ideation in Young Adolescents CAROL Z. GARRISON, PH.D., CHERYL L. ADDY, PH.D., KIRBY L. JACKSON, A.B., ROBERT E. McKEOWN, PH.D., AND JENNIFER L. WALLER, B.S.
Abstract. As part of a longitudinal study of depression and suicidal ideation in adolescents, a three-item suicide screen was administered to 1,073 students for 3 consecutive years starting at the beginning of the 7th or 8th grades. Each year over 70% of respondents reported no suicidal thoughts, and less than 5.5% attained high suicide ideation scores. Blacks and females had higher scores, respectively, than did whites and males. The individual students' suicide scores were less stable than the overall distributions with 1- and 2-year correlations reaching 0.35 and 0.28, respectively. Only one student received a high score all 3 years. The best predictors of a given year's suicide score was the previous years' depression scores. Gender, undesirable life events, family adaptability, and family cohesion were significant but less consistent predictors. J. Am. Acad. Child Adolesc. Psychiatry, 1991, 30, 4:597-603. Key Words: adolescence, community psychiatry, suicide, psychiatric status rating scales. Recently there has been much interest in identifying the frequency, correlates, and predictors of suicidal ideation in adolescence. Existing studies have tended to focus on clinical populations with only a handful addressing the problem in community samples of adolescents (Dubow et al., 1989; Harkavy-Friedman et al., 1987; Joffe et al., 1988; Kashani et aI., 1989; Levy and Deykin, 1989; Pfeffer et aI., 1984; Pronovost et al., 1990; Velez and Cohen, 1988). Findings from these community studies have not been uniform, with frequency estimates varying considerably from a low of 2% to a high of 60%. The observed variation in reported rates may in part reflect the demographic diversity of the populations investigated as well as the wide range of data collection methods employed (everything from anonymous selfadministered questionnaires to· clinician completed interviews). Findings must also be interpreted in light of the methodological constraints evidenced by some of the studies. Response rates have often been low, samples have been small and or select, and volunteers (i.e., those present at school on the day of data collection) were used. Within this diversity, a certain amount of constancy exists for, although a number of different correlates and predictors have been suggested, a recurring theme has been that depression, undesirable life events, and family environment may play important roles in the genesis of suicidal behaviors. Because none of the previous studies have reported on longitudinal results in community populations, the issue of predictors of suicidal ideation versus cross-sectional correlates has been left unaddressed. Accordingly, this paper reports on (I) the frequency, distribution, and stability of self-reported suicidal thoughts in a community sample of young adolescents
over a 3-year time period and (2) the relationship between depressive symptoms, undesirable life events, and perceived family environment and the onset or existence of suicidal thoughts in this age group.
Method The data for this investigation were collected during the screening phase of a larger longitudinal study of depressive symptomatology, major depressive disorder, and suicidal ideation in young adolescents. Data collection occurred annually in the fall of 3 consecutive years (1986-1988) from subjects attending four public middle schools and two high schools within a selected suburban school district in the southeast. All students enrolled in the 7th and 8th grades in 1986 were eligible for inclusion in the study. The data were collected in the classroom where students completed a self-administered questionnaire that included a demographic section, the 20-item Center for Epidemiologic Studies Depression Scale (CES-D) to which three suicide items were added, a modified version of the Coddington Life Events Scale for Adolescents (Coddington, 1972), and the 30-item Family Adaptability and Cohesion Evaluation Scales. The CES-D is a 20-item self-report symptom rating scale developed to measure depressive symptomatology in community adult populations (Radloff, 1977). The items included in the scale represent the major symptoms of the syndrome of depression as identified by clinical judgment, frequency of use in other questionnaires for depression, and factor analytic studies. When completing the CES-D, the subject is asked to report on his/her feelings during the preceding week. Responses are made on a four-point scale ("rarely or none of the time," "some or little of the time,"
Accepted October 2, 1990. From the Department of Epidemiology and Biostatistics, The University of South Carolina. This research was supported by Grant MH-40363 from the National Institute of Mental Health. Reprint requests to Dr. Garrison, Department of Epidemiology and Biostatistics, University ofSouth Carolina, Columbia, South Carolina 29208. 0890-8567/9113004-0597$03.00/0© 1991 by the American Academy of Child and Adolescent Psychiatry. J.Am.Acad. Child Adolesc.Psychiatry, 30:4, July 1991
"a lot of the time," and "most or all of the time") designed to measure the duration or frequency of symptoms. Each item has a possible value of zero to three; thus, the total score has a range of zero to 60. This single score is used to index the degree of depressive symptomatology present. In adult populations, scores of 16 or more generally have been used to indicate probable cases (Radloff, 1977). Although screening cutpoints for young adolescents have not yet been established, unpublished data with older adolescents suggest cutpoints of 22 in males and 24 in females
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provide the best combined sensitivity (84%) and specificity (75%) (Roberts et aI., 1990). The CES-D has been used in three studies of communitybased samples of young adolescents (Garrison et aI., 1989, 1990b; Schoenbach et aI., 1983). In all three investigations, the adolescents reported a higher proportion of transient symptoms in a comparison with previous fmdings with adults but similar (low) rates of persistent symptoms. These results raise questions regarding the comparability of total CES-D scores among different age groups. Garrison et a1. (1990c, unpublished manuscript) have reported on the internal consistency (coefficient alpha = 0.87) and on the 3-month stability of adolescent responses on the CES-D. The latter was assessed by individual item and total score comparisons. Only for one item (felt depressed) did the weighted kappa reach 0.40, the level regarded as "moderate agreement" (Fleiss, 1981). For other items, weighted kappas ranged from 0.01 to 0.38. The overall Spearman rank order correlation coefficient was moderately high (0.61 ,p < 0.000 1). The three suicide items added to the CES-D were "I felt that life was not worth living, " "I felt like hurting myself," and "I felt like killing myself." The total suicide score was obtained by summing responses on these three items, with a possible range of 0 to 9. A dichotomous suicide score was also computed with values greater than 5 indicating a high score. The Coddington Life Events Scale for Adolescents is a life events schedule that evolved out of the Holmes and Rahe (1967) approach to measuring life stress. When completing the instrument, the adolescent is asked to report the number of times within the past 12 months that he or she has experienced certain life events. The items are oriented toward the types of stressful life events relevant to the adolescent age group. The Coddington Scale has been modified for use in this study. Items involving drug use and pregnancy were deleted (at the request of the school district). Additional items including items involving college plans, employment, marriage, and driving were also omitted because of the peripheral nature of these concerns to most early adolescents. Two items dealing with physical appearance were added. It was thought that for the purpose of the present study, this area was not adequately addressed in the original scale. The modified life events scale contained a total of 41 events (19 desirable and 22 undesirable). (The designation of events as either desirable or undesirable was done with the realization that just about any event may be viewed as positive or negative depending on the circumstances surrounding the event and the way the event is perceived by the child.) Events could be reported as occurring more than once during the preceding year. An undesirable life events score was calculated as a simple unweighted sum of the total number of undesirable events reported. Items with missing responses were assigned a value of zero. The maximal possible score was 90. Unweighted rather than weighted sums were used, as the appropriateness of the weights assigned has not been well documented. In any case, available evidence suggests that weighted and unweighted scores are highly correlated with one another and similarly related to dependent measures
598
(Mueller et aI., 1977; Swearingen and Cohen, 1985; Tausig, 1982; Vinokur and Selzer, 1975). Family environment encompassed the adolescent's perception of the social and emotional climate within his family and was measured using the Family Adaptability and Cohesion Evaluation Scales (FACES II) (Olson et aI., 1982). FACES II is a 30-item, self-report written questionnaire that includes 16 cohesion items and 14 adaptability items. Cohesion items measured the emotional bonding that family members had toward one another as well as the individual autonomy experienced within the family system. Adaptability items measured the ability of the family system to change its power structure, role relationships, and rules in response to situational and developmental stress. When completing the questionnaire, individuals were asked to respond to each item on a five-point scale, ranging from "almost never true" to "almost always true." The possible range of scores for the family cohesion dimension was 16 to 80. A high score (greater than 64) denotes enmeshment and indicates that the family member perceives extreme closeness and limited autonomy in the family. A low score (less than 48) denotes disengagement and indicates a perception of low emotional bonding and high individual autonomy. A midrange score (48 to 64) indicates a balance between bonding and autonomy. The possible range for the total adaptability score was 14 to 70. A high score (greater than 52) denotes overwhelming change and characterizes the family as chaotically organized with capriciously shifting rules, roles, and power structure. A low score (less than 38) denotes little or no change and characterizes the family as rigidly organized. A moderate score (38 to 52) characterizes the family as having a balance between stability and change. FACES II evidences acceptable internal consistency, with alpha reliabilities for cohesion and adaptability of 0.88 and 0.78, respectively. Content validity has been shown by a confirmatory factor analysis that demonstrated independence of the two major factors. Construct validity has been supported in that the tool discriminates between clinic and nonclinic families. Additionally, data from a normative sample of 416 families with adolescents (1,248 individuals) are available (Olson et aI., 1982).
Results The initial sample included 1,083 individuals. Ten subjects in the "other" race (not white or black) category were excluded from the analyses. The remaining sample included the 1,073 individuals who participated in all 3 years of the study. Eight-five percent were white and 52% were female. At the study's onset, subjects ranged in age from 11 to 15 years, with the majority (99%) being between the ages of 12 and 14. Approximately 58% of the subjects lived in a family unit where both natural parents were present, while 20% lived in a single parent family, and 15% lived with one natural and one step-parent. A large number of subjects did not know the educational status of their mother (14%) or father (23%). Thirteen percent of mothers and 12% of fathers did not finish high school. Slightly more (17% of l.Am.Acad. Child Adolesc. Psychiatry, 30:4,July 1991
SUICIDAL IDEATION IN ADOLESCENCE
TABLE 1. Demographic Characteristics of Study Subjects (N = 1,073) Race Gender White males White females Black males Black females Age = 15 Guardian status Both natural parents Single parent I natural and 1 stepparent Other Education status of father Unknown Did not finish high school High school Technical school College graduate Educational status of mother Unknown Did not finish high school High school Technical school College graduate
N
%
444 467 71 91
41.4 43.5 6.6 8.5
420 508 132 13
39.1 47.3 12,3
623 219 157 74
58.1 20,4 14.6 6.9
391 101 236 136 209
36.4 9.4 22.0 12.7 19.5
282 118 372 123 178
26.3 11.0 34.7 11.5 16.6
1.2
the mothers and 22% of the fathers) were college graduates (Table I). The 1,073 participating students represent 70% of the 1,525 7th and 8th graders who completed the initial questionnaire in 1986. Since 95 to 98% of the students enrolled in the schools completed the questionnaire in any given year, the observed attrition (30%) primarily represents students who moved out of the school district and were no longer available for inclusion in the study. Comparison of those individuals who participated all 3 years with individuals participating less than 3 years indicates that latter individuals were somewhat more likely to be male (54% vs. 48%, p = 0.04), were older (13.07 vs. 12.75 years, p < 0.0001), were less likely to be living with both natural parents (39% vs. 58%, p < 0.0001), had higher CES-D (15.93 vs. 14.73, p = 0.03) and undesirable life events (9.83 vs. 7.32, p < 0.0001) scores, and had lower adaptability (43.73 vs. 45.81, P = 0.002) and cohesion (54.38 vs. 57.36, p < 0.0001) scores. The overall distribution of suicide scores was remarkably consistent for the 3-year period. In the cross-sectional data, each year at least 70% of the respondents reported no suicidal thoughts. Between 15 and 20% reported scores between 1 and 3, while less than 5.5% reported high scores greater than 5 (see Table 2). Comparisons indicated a significantly greater proportion of subjects with high suicide scores among females, for the total group, and within both the black and white subgroups (Table 3). The proportion of subjects with high suicide scores did not vary by race. A J.Am.Acad. Child Adolesc. Psychiatry, 30:4, July 1991
similar comparison of mean suicide scores for the various race-gender groups indicates that females consistently had higher scores than males and that blacks had higher scores than whites (Table 3). There was greater variability in individual students' suicide scores from year to year than in the overall distributions, with I and 2 year correlations reaching only 0.35 (1986-1987) and 0.28 (1986-1988), respectively (Table 4). A time related decrease is apparent in both the unadjusted (simple Pearson) correlations and in the residual correlations (0.34 for 1986-1987 and 0.26 for 1986-1988) that adjust for race and gender. Consideration of these same correlations in the specific race-gender groups shows higher correlations among blacks versus whites. Logistic regression analyses were done to explore the relationship of the various sociodemographic, life event, family, and depression scores to the dichotomized suicide score. Separate analyses were performed for each year and for the cross-sectional and longitudinal data. In a first set of longitudinal analyses, 1986 variables were used to predict the 1988 suicide score. In a second set, 1987 variables were used as predictors. Variables considered for entry into multivariable models included cohesion, adaptability, undesirable life events, CES-D score, and guardian status. Race, gender, and any other variable meeting a 0.05 significance level were retained in models. For ease of presentation, results for continuous variables (life events, cohesion, adaptability, and CES-D scores) are presented in terms of odds ratios standardized to represent a five-unit change in the variable. Cross-sectional results indicated that when considered in simple models controlling for race and gender, the undesirable life events, cohesion, adaptability, and CES-D scores were significantly associated (p < 0.05) with a high suicide score. In more complex stepwise multivariable models, which simultanously considered race, gender, and all variables that had been significant in the simple models, only the CES-D score remained significant all 3 years (odds ratios [OR] = 2.07 in 1986,2.38 in 1987, and 2.11 in 1988). Undesirable life events were significantly associated with high suicidal ideation in 1986 (OR = 1.42) and 1988 (OR = 1.44), as was cohesion in 1987 (OR = 0.68) and 1988 (OR = 0.73) and guardian status in 1987 (OR = 0.38) (Table 5). The longitudinal results were similar in several aspects to the cross-sectional findings. First, when considered separately, the previous years undesirable life events, cohesion, adaptability, and CES-D scores were significant predictors (p < 0.05) of later high suicide ideation. Second, in the more complex multivariable models both 1986 and 1987 CES-D scores were significant predictors of high suicidal ideation in 1988. Gender was another significant predictor in both years, as were the 1986 cohesion (OR = 0.83) and the 1987 adaptability (OR = 0.86) scores (Table 5). A repeated measures analysis of the dichotomized suicide scores was undertaken to account for the dependence of scores across the 3 years. The procedure for categorical data modeling (CATMOD) in the SAS User's Guide (1985) was used. This analysis revealed the same characteristics as the cross-sectional logistic regression analyses as well as several
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GARRISON ET AL. TABLE
2. Categorical Suicide Screening Scores by Year
Three-Year Cross-classifications
1988 Category
1986
1987
Category
Category
Zero
Zero Low Medium High Zero Low Medium High Zero Low Medium High Zero Low Medium High
Low
(1-3)
Medium
(4-5)
High (> = 6)
Zero
Low
High
Medium
N
%a
N
%"
N
%a
N
%a
557 45 6 5 104 24 5 4 12 2 2 2 7 7 2 2
51.91 4.19 0.56 0.47 9.69 2.24 0.47 0.37 1.12 0.19 0.19 0.19 0.65 0.65 0.19 0.19
68 28 6 3 25 21 4 2 6 5 I 2 4 2 2 0
6.34 2.61 0.56 0.28 2.33 1.96 0.37 0.19 0.56 0.47 0.09 0.19 0.37 0.19 0.19 0.00
9 3 2 5 4 4 2 2 2
0.84 0.28 0.19 0.47 0.37 0.37 0.09 0.19 0.19 0.19
3 2 5 I 5
0.28 0.19 0.47 0.09 0.47
17 4 3 2 5 7 2 4 2 2 2 1 2 1 3 I
1.58 0.37 0.28 0.19 0.47 0.65 0.19 0.37 0.19 0.19 0.19 0.09 0.19 0.09 0.28 0.09
46 42 50
4.29 3.91 4.66
46 43 58
4.29 4.01 5.41
I
Annual Summary Classifications
1986 1987 1988
763 826 786
71.11 76.98 73.25
218 162 179
20.32 15.10 16.68
apercentage of total, 1,073 students.
TABLE
3. Mean Suicide Scores, Standard Deviations (SD), and Percentage of Subjects with High Suicide Scores
(~6)
by Race-Gender Group and Year Suicide Score
1986 Race
Gender
N
Mean
SO
White
Male Female Male Female
444 467 71 91 1073
0.57 1.00 0.68 1.41 0.84
1.39 2.02 1.37 2.04 1.76
Black All
1987 Score
~6%
2.25 6.42 1.41 5.49 4.29
trends in the distribution of the suicide scores. As seen previously, race had no impact on the suicide scores all 3 years (p = 0.82). Gender, on the other hand, had a statistically significant impact on the distributions (p < 0.0001), with the proportion of females having high suicide scores 2.1 to 3.6 times higher than males; the effect of gender is essentially the same all 3 years (p = 0.47). The proportion of students with a high suicide score increased from year to year in every race-gender group except white females; this trend was of borderline statistical significance in black females (p = 0.06). To explore further the behavior of suicide scores through time, the scores were categorized as zero (0), low (1-3), medium (4-5), or high (6 and above). The frequency distributions of the categorized suicide scores for all 3 years are shown in Table 2. Transition probabilities from 1986 to 1988 and 1987 to 1988 were calculated for the total longitudinal sample (N = 1,073) and for the four race-gender 600
Mean
SO
0.47 0.81 0.73 1.58 0.73
1.47 1.87 1.65 2.30 1.77
1988 Score
~6%
2.70 4.71 1.41 8.79 4.01
Mean
SO
0.57 1.19 0.66 1.52 0.92
1.52 2.19 1.76 2.48 1.97
Score
~6%
2.93 7.07 2.82 10.99 5.41
groups. The transition probabilities for the total sample are summarized in Table 6; the probabilities behave similarly in the four race-gender groups. For example, in the total sample there are 218 students who have a low (1-3) suicide score in 1986 (summing all frequencies in rows 5-8 in Table 2); 137 of these students have a zero score in 1988 (summing all frequencies in first column of rows 5-8), indicating a 62.8% (1371218) probability of obtaining a score of zero in 1988, given a low score in 1986. In the total sample and in the four race-gender groups, these transition probabilities reflect the marginal distributions of the suicide scores, in that transition into the zero category dominates transition into any other category. However, the higher the earlier score, the lower the probability of having a zero score in 1988 (p < 0.0001 for both 1- and 2-year transitions). The reverse is seen in the high score category: the, probability of having a transition into this category is typically lower than the other possible transiJ. Am .Acad. Child Adolesc. Psychiatry, 30:4, July J99 J
SUICIDAL IDEATION IN ADOLESCENCE TABLE
4. Pearson Correlation of Suicide Screening Scores over Time
TABLE
Race-Gender
1986-1987
1986-1988
White males White females Black males Black females Total
0.22 0.36 0.44 0.44 0.35
0.20 0.27 0.36 0.32 0.28
1988 Suicide Score 1986 suicide score 0 1-3 4--5 6+ 1987 suicide score 0 1-3 4--5 6+
tions, but the higher the earlier suicide score, the higher the probability of having a high score in 1988 (p < 0.0001 for both 1- and 2-year transitions). This observation is tempered by noting that there is little distinction between the medium (score = 4-5) and high (score ;:::6) categories in terms of the transition probabilities. It is of interest that only one individual, a white female, had high suicide scores all 3 years.
The overall distribution of suicide scores was remarkably consistent over the 3-year period, with nearly three-quarters of subjects reporting no suicidal ideation, and 5.5% or less reporting high levels of ideation during the previous week. These estimates are considerably lower than those of Smith and Crawford (1986) who found 63% of a sample of midwestern high school students reported some ideation and Harkavy-Friedman et al. (1987) who found 60% of an academically select New York high school sample reported ideation. However, the current findings are not dissimilar from those of Dubow et al. (1989) who found 7.7% of a sample of junior high and high school students in a semirural community reported extremely troubling suicidal ideation; Kashani et al. (1989) who found 6.6% of a sample of 8-, 12-, and 17-year-olds reported recurrent suicidal ideation; Velez and Cohen (1988) who found 6.7% of a New York sample of 9- to 18-year-olds reported current suicidal ideation; and Joffe et al. (1988) who found 5 to 10% of
1-3
4--5
6+
Total Frequency
80.3" 62.8 39.1 39.1
13.8 23.9 30.4 17.4
2.5 5.0 15.2 28.3
3.4 8.3 15.2 15.2
763 218 46 46
82.3 48.1 35.7 30.2
12.5 34.6 31.0 16.3
2.1 8.6 9.5 34.9
3.1 8.6 23.8 18.6
826 162 42 43
male and 10 to 20% of female youth participating in the Ontario Child Health Survey reported suicidal behavior within a six-month period. The differences in the rates observed may be the result of several factors including the following: (1) variation in the demographic profile (age, race, gender, SES, and geographic location) of the populations sampled; (2) variation in the time frame assessed (current, last week, past 6 months, past year, and lifetime); and (3) variation in the nature of the questions asked (general thoughts about death versus specific and persistent suicidal thoughts and plans) and the manner in which suicidal ideation is defined. Velez and Cohen (1988) have suggested and shown that the prevalence of suicidal ideation in adolescent populations will vary greatly depending on how stringently such ideation is defined. However the present results support the conclusions of Kashani et al. (1989) who previously indicated that even when applying a fairly stringent definition of suicidal ideation, the proportion of adolescents reporting serious suicidal thoughts is large enough to be of concern. The finding that females attain higher suicide scores than
5. Cross-Sectional and Longitudinal Multivariable Logistic Regression Models of Suicide Scores with Depression, Life Event, Family, and Demographic Variables
Variable Cross-sectional model Race Gender Guardian statusb Undesirable life events" Cohesion'
CES-IY
o
aOf those who had a zero suicide score in 1986, 80.3% had a zero score in 1988.
Discussion
TABLE
6. Transition Probabilities (%)for Categorized Suicide Score for Total Longitudinal Sample (N=1,073)
Odds Ratio 95% CIa
Odds Ratio 95% CIa
Odds Ratio 95% CIa
1986 0.64 (0.23, 1.76) 1.47 (0.65, 3.34)
1987 1.24 (0.44, 3.53) 0.59 (0.23, 1.53) 0.38 (0.16,0.90)
1988 1.53 (0.64, 3.68) 0.64 (0.27, 1.54)
0.68 (0.56, 0.82)
0.73 (0.63,0.86)
1.42 (1.14, 1.77) 2.07 (1.72, 2.50)
Longitudinal model Race Gender Cohesion' Adaptability'
CES-IY
1.44 (1.18, 1.77) 2.38 (1.91,2.97) 1986 predicting 1988 1987 predicting 1988 1.22 (0.61, 2.44) 1.23 (0.61,2.47) 1.98 (1.06,3.71) 1.97 (1.05,3.69) 0.83 (0.74,0.94) 0.86 (0.75, 0.98) 1.19 (1.03, 1.37) 1.30 (1.14, 1.48)
2.11 (1.73, 2.56)
"Confidence interval. bThe reference category is living with both natural parents. "Odds ratio for 5 unit change in score.
l.Am.Acad. Child Adolesc.Psychiany, 30:4,luly 1991
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GARRISON ET AL.
males is consistent with a number of previous adolescent studies (Dubow et aI., 1989; Joffe et aI., 1989; Kashani et aI., 1989; Pronovost et aI., 1990), although not all studies have found this gender differential (Harkavy-Friedman et al., 1987). Of interest is the fact that this gender difference remains, though somewhat attenuated, even when the level of depressive symptomatology reported is taken into account. Thus, the higher rates of suicidal ideation do not seem to merely reflect higher rates of depressive symptomatology among females. The higher female ideation rates coincide with higher rates of suicide attempts observed in adolescent and adult females (Moscicki et al., 1988) but do not correspond to completed suicides where rates in males exceed those in females (Monk, 1987). Future comparisons of the composition and persistence of suicidal ideation in males and females might improve our understanding of the observed gender differences and our ability to predict outcomes associated with varying constellations of suicidal behavior. Findings suggest that adolescent suicide scores vary considerably over time. Although 90% of the low scorers (scores
Abstract. As part of a longitudinal study of depression and suicidal ideation in adolescents, a three-item suicide screen was administered to 1,073 students for 3 consecutive years starting at the beginning of the 7th or 8th grades. Each year over 70% of respondents reported no suicidal thoughts, and less than 5.5% attained high suicide ideation scores. Blacks and females had higher scores, respectively, than did whites and males. The individual students' suicide scores were less stable than the overall distributions with 1- and 2-year correlations reaching 0.35 and 0.28, respectively. Only one student received a high score all 3 years. The best predictors of a given year's suicide score was the previous years' depression scores. Gender, undesirable life events, family adaptability, and family cohesion were significant but less consistent predictors. J. Am. Acad. Child Adolesc. Psychiatry, 1991, 30, 4:597-603. Key Words: adolescence, community psychiatry, suicide, psychiatric status rating scales. Recently there has been much interest in identifying the frequency, correlates, and predictors of suicidal ideation in adolescence. Existing studies have tended to focus on clinical populations with only a handful addressing the problem in community samples of adolescents (Dubow et al., 1989; Harkavy-Friedman et al., 1987; Joffe et al., 1988; Kashani et aI., 1989; Levy and Deykin, 1989; Pfeffer et aI., 1984; Pronovost et al., 1990; Velez and Cohen, 1988). Findings from these community studies have not been uniform, with frequency estimates varying considerably from a low of 2% to a high of 60%. The observed variation in reported rates may in part reflect the demographic diversity of the populations investigated as well as the wide range of data collection methods employed (everything from anonymous selfadministered questionnaires to· clinician completed interviews). Findings must also be interpreted in light of the methodological constraints evidenced by some of the studies. Response rates have often been low, samples have been small and or select, and volunteers (i.e., those present at school on the day of data collection) were used. Within this diversity, a certain amount of constancy exists for, although a number of different correlates and predictors have been suggested, a recurring theme has been that depression, undesirable life events, and family environment may play important roles in the genesis of suicidal behaviors. Because none of the previous studies have reported on longitudinal results in community populations, the issue of predictors of suicidal ideation versus cross-sectional correlates has been left unaddressed. Accordingly, this paper reports on (I) the frequency, distribution, and stability of self-reported suicidal thoughts in a community sample of young adolescents
over a 3-year time period and (2) the relationship between depressive symptoms, undesirable life events, and perceived family environment and the onset or existence of suicidal thoughts in this age group.
Method The data for this investigation were collected during the screening phase of a larger longitudinal study of depressive symptomatology, major depressive disorder, and suicidal ideation in young adolescents. Data collection occurred annually in the fall of 3 consecutive years (1986-1988) from subjects attending four public middle schools and two high schools within a selected suburban school district in the southeast. All students enrolled in the 7th and 8th grades in 1986 were eligible for inclusion in the study. The data were collected in the classroom where students completed a self-administered questionnaire that included a demographic section, the 20-item Center for Epidemiologic Studies Depression Scale (CES-D) to which three suicide items were added, a modified version of the Coddington Life Events Scale for Adolescents (Coddington, 1972), and the 30-item Family Adaptability and Cohesion Evaluation Scales. The CES-D is a 20-item self-report symptom rating scale developed to measure depressive symptomatology in community adult populations (Radloff, 1977). The items included in the scale represent the major symptoms of the syndrome of depression as identified by clinical judgment, frequency of use in other questionnaires for depression, and factor analytic studies. When completing the CES-D, the subject is asked to report on his/her feelings during the preceding week. Responses are made on a four-point scale ("rarely or none of the time," "some or little of the time,"
Accepted October 2, 1990. From the Department of Epidemiology and Biostatistics, The University of South Carolina. This research was supported by Grant MH-40363 from the National Institute of Mental Health. Reprint requests to Dr. Garrison, Department of Epidemiology and Biostatistics, University ofSouth Carolina, Columbia, South Carolina 29208. 0890-8567/9113004-0597$03.00/0© 1991 by the American Academy of Child and Adolescent Psychiatry. J.Am.Acad. Child Adolesc.Psychiatry, 30:4, July 1991
"a lot of the time," and "most or all of the time") designed to measure the duration or frequency of symptoms. Each item has a possible value of zero to three; thus, the total score has a range of zero to 60. This single score is used to index the degree of depressive symptomatology present. In adult populations, scores of 16 or more generally have been used to indicate probable cases (Radloff, 1977). Although screening cutpoints for young adolescents have not yet been established, unpublished data with older adolescents suggest cutpoints of 22 in males and 24 in females
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GARRISON ET AL.
provide the best combined sensitivity (84%) and specificity (75%) (Roberts et aI., 1990). The CES-D has been used in three studies of communitybased samples of young adolescents (Garrison et aI., 1989, 1990b; Schoenbach et aI., 1983). In all three investigations, the adolescents reported a higher proportion of transient symptoms in a comparison with previous fmdings with adults but similar (low) rates of persistent symptoms. These results raise questions regarding the comparability of total CES-D scores among different age groups. Garrison et a1. (1990c, unpublished manuscript) have reported on the internal consistency (coefficient alpha = 0.87) and on the 3-month stability of adolescent responses on the CES-D. The latter was assessed by individual item and total score comparisons. Only for one item (felt depressed) did the weighted kappa reach 0.40, the level regarded as "moderate agreement" (Fleiss, 1981). For other items, weighted kappas ranged from 0.01 to 0.38. The overall Spearman rank order correlation coefficient was moderately high (0.61 ,p < 0.000 1). The three suicide items added to the CES-D were "I felt that life was not worth living, " "I felt like hurting myself," and "I felt like killing myself." The total suicide score was obtained by summing responses on these three items, with a possible range of 0 to 9. A dichotomous suicide score was also computed with values greater than 5 indicating a high score. The Coddington Life Events Scale for Adolescents is a life events schedule that evolved out of the Holmes and Rahe (1967) approach to measuring life stress. When completing the instrument, the adolescent is asked to report the number of times within the past 12 months that he or she has experienced certain life events. The items are oriented toward the types of stressful life events relevant to the adolescent age group. The Coddington Scale has been modified for use in this study. Items involving drug use and pregnancy were deleted (at the request of the school district). Additional items including items involving college plans, employment, marriage, and driving were also omitted because of the peripheral nature of these concerns to most early adolescents. Two items dealing with physical appearance were added. It was thought that for the purpose of the present study, this area was not adequately addressed in the original scale. The modified life events scale contained a total of 41 events (19 desirable and 22 undesirable). (The designation of events as either desirable or undesirable was done with the realization that just about any event may be viewed as positive or negative depending on the circumstances surrounding the event and the way the event is perceived by the child.) Events could be reported as occurring more than once during the preceding year. An undesirable life events score was calculated as a simple unweighted sum of the total number of undesirable events reported. Items with missing responses were assigned a value of zero. The maximal possible score was 90. Unweighted rather than weighted sums were used, as the appropriateness of the weights assigned has not been well documented. In any case, available evidence suggests that weighted and unweighted scores are highly correlated with one another and similarly related to dependent measures
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(Mueller et aI., 1977; Swearingen and Cohen, 1985; Tausig, 1982; Vinokur and Selzer, 1975). Family environment encompassed the adolescent's perception of the social and emotional climate within his family and was measured using the Family Adaptability and Cohesion Evaluation Scales (FACES II) (Olson et aI., 1982). FACES II is a 30-item, self-report written questionnaire that includes 16 cohesion items and 14 adaptability items. Cohesion items measured the emotional bonding that family members had toward one another as well as the individual autonomy experienced within the family system. Adaptability items measured the ability of the family system to change its power structure, role relationships, and rules in response to situational and developmental stress. When completing the questionnaire, individuals were asked to respond to each item on a five-point scale, ranging from "almost never true" to "almost always true." The possible range of scores for the family cohesion dimension was 16 to 80. A high score (greater than 64) denotes enmeshment and indicates that the family member perceives extreme closeness and limited autonomy in the family. A low score (less than 48) denotes disengagement and indicates a perception of low emotional bonding and high individual autonomy. A midrange score (48 to 64) indicates a balance between bonding and autonomy. The possible range for the total adaptability score was 14 to 70. A high score (greater than 52) denotes overwhelming change and characterizes the family as chaotically organized with capriciously shifting rules, roles, and power structure. A low score (less than 38) denotes little or no change and characterizes the family as rigidly organized. A moderate score (38 to 52) characterizes the family as having a balance between stability and change. FACES II evidences acceptable internal consistency, with alpha reliabilities for cohesion and adaptability of 0.88 and 0.78, respectively. Content validity has been shown by a confirmatory factor analysis that demonstrated independence of the two major factors. Construct validity has been supported in that the tool discriminates between clinic and nonclinic families. Additionally, data from a normative sample of 416 families with adolescents (1,248 individuals) are available (Olson et aI., 1982).
Results The initial sample included 1,083 individuals. Ten subjects in the "other" race (not white or black) category were excluded from the analyses. The remaining sample included the 1,073 individuals who participated in all 3 years of the study. Eight-five percent were white and 52% were female. At the study's onset, subjects ranged in age from 11 to 15 years, with the majority (99%) being between the ages of 12 and 14. Approximately 58% of the subjects lived in a family unit where both natural parents were present, while 20% lived in a single parent family, and 15% lived with one natural and one step-parent. A large number of subjects did not know the educational status of their mother (14%) or father (23%). Thirteen percent of mothers and 12% of fathers did not finish high school. Slightly more (17% of l.Am.Acad. Child Adolesc. Psychiatry, 30:4,July 1991
SUICIDAL IDEATION IN ADOLESCENCE
TABLE 1. Demographic Characteristics of Study Subjects (N = 1,073) Race Gender White males White females Black males Black females Age = 15 Guardian status Both natural parents Single parent I natural and 1 stepparent Other Education status of father Unknown Did not finish high school High school Technical school College graduate Educational status of mother Unknown Did not finish high school High school Technical school College graduate
N
%
444 467 71 91
41.4 43.5 6.6 8.5
420 508 132 13
39.1 47.3 12,3
623 219 157 74
58.1 20,4 14.6 6.9
391 101 236 136 209
36.4 9.4 22.0 12.7 19.5
282 118 372 123 178
26.3 11.0 34.7 11.5 16.6
1.2
the mothers and 22% of the fathers) were college graduates (Table I). The 1,073 participating students represent 70% of the 1,525 7th and 8th graders who completed the initial questionnaire in 1986. Since 95 to 98% of the students enrolled in the schools completed the questionnaire in any given year, the observed attrition (30%) primarily represents students who moved out of the school district and were no longer available for inclusion in the study. Comparison of those individuals who participated all 3 years with individuals participating less than 3 years indicates that latter individuals were somewhat more likely to be male (54% vs. 48%, p = 0.04), were older (13.07 vs. 12.75 years, p < 0.0001), were less likely to be living with both natural parents (39% vs. 58%, p < 0.0001), had higher CES-D (15.93 vs. 14.73, p = 0.03) and undesirable life events (9.83 vs. 7.32, p < 0.0001) scores, and had lower adaptability (43.73 vs. 45.81, P = 0.002) and cohesion (54.38 vs. 57.36, p < 0.0001) scores. The overall distribution of suicide scores was remarkably consistent for the 3-year period. In the cross-sectional data, each year at least 70% of the respondents reported no suicidal thoughts. Between 15 and 20% reported scores between 1 and 3, while less than 5.5% reported high scores greater than 5 (see Table 2). Comparisons indicated a significantly greater proportion of subjects with high suicide scores among females, for the total group, and within both the black and white subgroups (Table 3). The proportion of subjects with high suicide scores did not vary by race. A J.Am.Acad. Child Adolesc. Psychiatry, 30:4, July 1991
similar comparison of mean suicide scores for the various race-gender groups indicates that females consistently had higher scores than males and that blacks had higher scores than whites (Table 3). There was greater variability in individual students' suicide scores from year to year than in the overall distributions, with I and 2 year correlations reaching only 0.35 (1986-1987) and 0.28 (1986-1988), respectively (Table 4). A time related decrease is apparent in both the unadjusted (simple Pearson) correlations and in the residual correlations (0.34 for 1986-1987 and 0.26 for 1986-1988) that adjust for race and gender. Consideration of these same correlations in the specific race-gender groups shows higher correlations among blacks versus whites. Logistic regression analyses were done to explore the relationship of the various sociodemographic, life event, family, and depression scores to the dichotomized suicide score. Separate analyses were performed for each year and for the cross-sectional and longitudinal data. In a first set of longitudinal analyses, 1986 variables were used to predict the 1988 suicide score. In a second set, 1987 variables were used as predictors. Variables considered for entry into multivariable models included cohesion, adaptability, undesirable life events, CES-D score, and guardian status. Race, gender, and any other variable meeting a 0.05 significance level were retained in models. For ease of presentation, results for continuous variables (life events, cohesion, adaptability, and CES-D scores) are presented in terms of odds ratios standardized to represent a five-unit change in the variable. Cross-sectional results indicated that when considered in simple models controlling for race and gender, the undesirable life events, cohesion, adaptability, and CES-D scores were significantly associated (p < 0.05) with a high suicide score. In more complex stepwise multivariable models, which simultanously considered race, gender, and all variables that had been significant in the simple models, only the CES-D score remained significant all 3 years (odds ratios [OR] = 2.07 in 1986,2.38 in 1987, and 2.11 in 1988). Undesirable life events were significantly associated with high suicidal ideation in 1986 (OR = 1.42) and 1988 (OR = 1.44), as was cohesion in 1987 (OR = 0.68) and 1988 (OR = 0.73) and guardian status in 1987 (OR = 0.38) (Table 5). The longitudinal results were similar in several aspects to the cross-sectional findings. First, when considered separately, the previous years undesirable life events, cohesion, adaptability, and CES-D scores were significant predictors (p < 0.05) of later high suicide ideation. Second, in the more complex multivariable models both 1986 and 1987 CES-D scores were significant predictors of high suicidal ideation in 1988. Gender was another significant predictor in both years, as were the 1986 cohesion (OR = 0.83) and the 1987 adaptability (OR = 0.86) scores (Table 5). A repeated measures analysis of the dichotomized suicide scores was undertaken to account for the dependence of scores across the 3 years. The procedure for categorical data modeling (CATMOD) in the SAS User's Guide (1985) was used. This analysis revealed the same characteristics as the cross-sectional logistic regression analyses as well as several
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GARRISON ET AL. TABLE
2. Categorical Suicide Screening Scores by Year
Three-Year Cross-classifications
1988 Category
1986
1987
Category
Category
Zero
Zero Low Medium High Zero Low Medium High Zero Low Medium High Zero Low Medium High
Low
(1-3)
Medium
(4-5)
High (> = 6)
Zero
Low
High
Medium
N
%a
N
%"
N
%a
N
%a
557 45 6 5 104 24 5 4 12 2 2 2 7 7 2 2
51.91 4.19 0.56 0.47 9.69 2.24 0.47 0.37 1.12 0.19 0.19 0.19 0.65 0.65 0.19 0.19
68 28 6 3 25 21 4 2 6 5 I 2 4 2 2 0
6.34 2.61 0.56 0.28 2.33 1.96 0.37 0.19 0.56 0.47 0.09 0.19 0.37 0.19 0.19 0.00
9 3 2 5 4 4 2 2 2
0.84 0.28 0.19 0.47 0.37 0.37 0.09 0.19 0.19 0.19
3 2 5 I 5
0.28 0.19 0.47 0.09 0.47
17 4 3 2 5 7 2 4 2 2 2 1 2 1 3 I
1.58 0.37 0.28 0.19 0.47 0.65 0.19 0.37 0.19 0.19 0.19 0.09 0.19 0.09 0.28 0.09
46 42 50
4.29 3.91 4.66
46 43 58
4.29 4.01 5.41
I
Annual Summary Classifications
1986 1987 1988
763 826 786
71.11 76.98 73.25
218 162 179
20.32 15.10 16.68
apercentage of total, 1,073 students.
TABLE
3. Mean Suicide Scores, Standard Deviations (SD), and Percentage of Subjects with High Suicide Scores
(~6)
by Race-Gender Group and Year Suicide Score
1986 Race
Gender
N
Mean
SO
White
Male Female Male Female
444 467 71 91 1073
0.57 1.00 0.68 1.41 0.84
1.39 2.02 1.37 2.04 1.76
Black All
1987 Score
~6%
2.25 6.42 1.41 5.49 4.29
trends in the distribution of the suicide scores. As seen previously, race had no impact on the suicide scores all 3 years (p = 0.82). Gender, on the other hand, had a statistically significant impact on the distributions (p < 0.0001), with the proportion of females having high suicide scores 2.1 to 3.6 times higher than males; the effect of gender is essentially the same all 3 years (p = 0.47). The proportion of students with a high suicide score increased from year to year in every race-gender group except white females; this trend was of borderline statistical significance in black females (p = 0.06). To explore further the behavior of suicide scores through time, the scores were categorized as zero (0), low (1-3), medium (4-5), or high (6 and above). The frequency distributions of the categorized suicide scores for all 3 years are shown in Table 2. Transition probabilities from 1986 to 1988 and 1987 to 1988 were calculated for the total longitudinal sample (N = 1,073) and for the four race-gender 600
Mean
SO
0.47 0.81 0.73 1.58 0.73
1.47 1.87 1.65 2.30 1.77
1988 Score
~6%
2.70 4.71 1.41 8.79 4.01
Mean
SO
0.57 1.19 0.66 1.52 0.92
1.52 2.19 1.76 2.48 1.97
Score
~6%
2.93 7.07 2.82 10.99 5.41
groups. The transition probabilities for the total sample are summarized in Table 6; the probabilities behave similarly in the four race-gender groups. For example, in the total sample there are 218 students who have a low (1-3) suicide score in 1986 (summing all frequencies in rows 5-8 in Table 2); 137 of these students have a zero score in 1988 (summing all frequencies in first column of rows 5-8), indicating a 62.8% (1371218) probability of obtaining a score of zero in 1988, given a low score in 1986. In the total sample and in the four race-gender groups, these transition probabilities reflect the marginal distributions of the suicide scores, in that transition into the zero category dominates transition into any other category. However, the higher the earlier score, the lower the probability of having a zero score in 1988 (p < 0.0001 for both 1- and 2-year transitions). The reverse is seen in the high score category: the, probability of having a transition into this category is typically lower than the other possible transiJ. Am .Acad. Child Adolesc. Psychiatry, 30:4, July J99 J
SUICIDAL IDEATION IN ADOLESCENCE TABLE
4. Pearson Correlation of Suicide Screening Scores over Time
TABLE
Race-Gender
1986-1987
1986-1988
White males White females Black males Black females Total
0.22 0.36 0.44 0.44 0.35
0.20 0.27 0.36 0.32 0.28
1988 Suicide Score 1986 suicide score 0 1-3 4--5 6+ 1987 suicide score 0 1-3 4--5 6+
tions, but the higher the earlier suicide score, the higher the probability of having a high score in 1988 (p < 0.0001 for both 1- and 2-year transitions). This observation is tempered by noting that there is little distinction between the medium (score = 4-5) and high (score ;:::6) categories in terms of the transition probabilities. It is of interest that only one individual, a white female, had high suicide scores all 3 years.
The overall distribution of suicide scores was remarkably consistent over the 3-year period, with nearly three-quarters of subjects reporting no suicidal ideation, and 5.5% or less reporting high levels of ideation during the previous week. These estimates are considerably lower than those of Smith and Crawford (1986) who found 63% of a sample of midwestern high school students reported some ideation and Harkavy-Friedman et al. (1987) who found 60% of an academically select New York high school sample reported ideation. However, the current findings are not dissimilar from those of Dubow et al. (1989) who found 7.7% of a sample of junior high and high school students in a semirural community reported extremely troubling suicidal ideation; Kashani et al. (1989) who found 6.6% of a sample of 8-, 12-, and 17-year-olds reported recurrent suicidal ideation; Velez and Cohen (1988) who found 6.7% of a New York sample of 9- to 18-year-olds reported current suicidal ideation; and Joffe et al. (1988) who found 5 to 10% of
1-3
4--5
6+
Total Frequency
80.3" 62.8 39.1 39.1
13.8 23.9 30.4 17.4
2.5 5.0 15.2 28.3
3.4 8.3 15.2 15.2
763 218 46 46
82.3 48.1 35.7 30.2
12.5 34.6 31.0 16.3
2.1 8.6 9.5 34.9
3.1 8.6 23.8 18.6
826 162 42 43
male and 10 to 20% of female youth participating in the Ontario Child Health Survey reported suicidal behavior within a six-month period. The differences in the rates observed may be the result of several factors including the following: (1) variation in the demographic profile (age, race, gender, SES, and geographic location) of the populations sampled; (2) variation in the time frame assessed (current, last week, past 6 months, past year, and lifetime); and (3) variation in the nature of the questions asked (general thoughts about death versus specific and persistent suicidal thoughts and plans) and the manner in which suicidal ideation is defined. Velez and Cohen (1988) have suggested and shown that the prevalence of suicidal ideation in adolescent populations will vary greatly depending on how stringently such ideation is defined. However the present results support the conclusions of Kashani et al. (1989) who previously indicated that even when applying a fairly stringent definition of suicidal ideation, the proportion of adolescents reporting serious suicidal thoughts is large enough to be of concern. The finding that females attain higher suicide scores than
5. Cross-Sectional and Longitudinal Multivariable Logistic Regression Models of Suicide Scores with Depression, Life Event, Family, and Demographic Variables
Variable Cross-sectional model Race Gender Guardian statusb Undesirable life events" Cohesion'
CES-IY
o
aOf those who had a zero suicide score in 1986, 80.3% had a zero score in 1988.
Discussion
TABLE
6. Transition Probabilities (%)for Categorized Suicide Score for Total Longitudinal Sample (N=1,073)
Odds Ratio 95% CIa
Odds Ratio 95% CIa
Odds Ratio 95% CIa
1986 0.64 (0.23, 1.76) 1.47 (0.65, 3.34)
1987 1.24 (0.44, 3.53) 0.59 (0.23, 1.53) 0.38 (0.16,0.90)
1988 1.53 (0.64, 3.68) 0.64 (0.27, 1.54)
0.68 (0.56, 0.82)
0.73 (0.63,0.86)
1.42 (1.14, 1.77) 2.07 (1.72, 2.50)
Longitudinal model Race Gender Cohesion' Adaptability'
CES-IY
1.44 (1.18, 1.77) 2.38 (1.91,2.97) 1986 predicting 1988 1987 predicting 1988 1.22 (0.61, 2.44) 1.23 (0.61,2.47) 1.98 (1.06,3.71) 1.97 (1.05,3.69) 0.83 (0.74,0.94) 0.86 (0.75, 0.98) 1.19 (1.03, 1.37) 1.30 (1.14, 1.48)
2.11 (1.73, 2.56)
"Confidence interval. bThe reference category is living with both natural parents. "Odds ratio for 5 unit change in score.
l.Am.Acad. Child Adolesc.Psychiany, 30:4,luly 1991
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GARRISON ET AL.
males is consistent with a number of previous adolescent studies (Dubow et aI., 1989; Joffe et aI., 1989; Kashani et aI., 1989; Pronovost et aI., 1990), although not all studies have found this gender differential (Harkavy-Friedman et al., 1987). Of interest is the fact that this gender difference remains, though somewhat attenuated, even when the level of depressive symptomatology reported is taken into account. Thus, the higher rates of suicidal ideation do not seem to merely reflect higher rates of depressive symptomatology among females. The higher female ideation rates coincide with higher rates of suicide attempts observed in adolescent and adult females (Moscicki et al., 1988) but do not correspond to completed suicides where rates in males exceed those in females (Monk, 1987). Future comparisons of the composition and persistence of suicidal ideation in males and females might improve our understanding of the observed gender differences and our ability to predict outcomes associated with varying constellations of suicidal behavior. Findings suggest that adolescent suicide scores vary considerably over time. Although 90% of the low scorers (scores
