In press, Aggressive Behavior
Unemployment and Coerced Treatment: A Test of the Intolerance
Hypothesis*
Ralph Catalano
Lonnie Snowden
University of California, Berkeley
Martha Shumway
University of California, San Francisco
Running Head: Economic Antecedents of Intolerance
Word Count: 3,747
Tables: 3
Figures: 0
Key Words: Unemployment, coerced treatment, tolerance
*NIMH grant P50 MH43694-12 supported the analyses described in this manuscript. Direct correspondence to Ralph Catalano, School of Public Health,
University of California, Berkeley, California 94720; electronic mail: rayc@uclink4.berkeley.edu.

Unemployment and Coerced Treatment: A Test of the Intolerance
Hypothesis*
Theory suggests that the reported association between economic indicators and the incidence of coerced treatment for mental illness may result, at least in part, from reduced tolerance in the community for impaired behavior among minorities.
We hypothesize that the median level of functioning among those coerced into treatment will vary positively with earlier incidence of unemployment. The test applies Box-Jenkins methods to 156 months (August 1985 through July 1998) of data from California. Consistent with theory, results support the hypothesis only for African American males.

1
Unemployment and Coerced Treatment: A Test of the Intolerance
Hypothesis*
The circumstances that lead to coerced treatment of mentally ill persons appear complex (Hiday 1992; Monahan et al. 1999). Despite this complexity, research finds that the incidence of such treatment varies over time in ways that suggest shared catalysts (Catalano et al. 2003). These catalysts apparently include rising unemployment.
Catalano, Novaco, and McConnell (1997; 2002) empirically measured the association between unemployment and the incidence of coerced treatment in
Pittsburgh, Pennsylvania and San Francisco, California. Both tests found that coerced treatment rises above otherwise expected levels after unemployment increased. The association appeared among men and women in San Francisco but only among men in Pittsburgh.
The larger literature reporting an association between rising unemployment and the use of inpatient mental health services typically attributes the relationship to mechanisms in which economic hard times increase the number of individuals who exhibit mental illness (Barling and Handal 1980; Bland et al. 1988. Brenner 1967;Catalano, Dooley, and Jackson 1985; Dowdall,
Marshall, and Morra 1990; Kiernan et al. 1989; Marshall and Dowdall 1982;
Marshall and Funch 1979; Stokes and Cochrane 1984). The mechanisms thought to connect economic contraction with disorder include stress (Dooley and
Catalano 1984), the disruption of services for the chronically mentally ill

2
(Rodriguez 2001), and aggressive behavior induced by economic frustration
(Catalano, Novaco, and McConnell 1997; 2002).
Epidemiologic research lends credence to the argument that the association between rising unemployment and coerced treatment may arise from mental illness and aggression induced by the economy. Alcohol abuse and antisocial behavior reportedly increases among persons who seek, but cannot find, work (Dee 2001; Dooley et al . 1992; Dooley and Prause 1998; Rodriguez et al . 2001). Data from the last true panel measurement of the psychiatric status of
Americans, for example, show that persons who, at first assessment, had no history of either diagnosable alcohol abuse or anti-social behavior, exhibited elevated risk of each if laid off before their second assessment (Catalano et al .
1993a; Catalano et al . 1993b). Losing a job multiplied the odds of diagnosable alcohol abuse by 5 and the odds of antisocial behavior by 6.
The theory and analyses described in this paper focus on an alternative, though related, explanation of the association between rising unemployment and coerced treatment. Drawing on theories from sociology and social psychology, we argue that some part of the association could be due to a circumstance in which economic contraction causes those in contact with noisome and behaviorally impaired persons to become less tolerant of such deviance.
Mental health authorities routinely coerce into treatment those persons they believe to be psychiatrically impaired and whose behavior has induced grave disability, or been judged suicidal or a threat to others (Monahan et al.
1995; Ross et al . 1996). Formal rules supposedly guide clinicians in determining

3 whether individuals meet the impairment, disability, and threat criteria for coerced treatment (Faulkner et al . 1986). Clinicians, however, cannot observe everyone whose behavior makes him or her a candidate for coerced treatment. They instead assess a sample presented by police, other first responders, or the family or friends of candidates. The characteristics of the sample necessarily reflect the biases and tolerance that the community at large brings to encounters with behaviorally impaired persons.
One mechanism that we believe may lead to reduced tolerance for persons who meet the criteria for coerced treatment arises from classic frustration aggression theory (Berkowitz 1989; Dollard et al. 1939). The theory posits, in essence, that persons become frustrated and or angry when they believe that circumstances unrelated to their behavior deny them an expected reward. They supposedly cope with these feelings by overt or covert aggression that relieves their frustration.
Researchers have, as noted above, cited frustration aggression theory as a source of the hypothesis that economic contraction causes more people to meet the criteria for coerced treatment. This argument assumes that a slowing economy causes a growing number of people to believe that circumstances unrelated to their performance have denied them deserved improvements in socioeconomic status. This belief induces frustration, and a fraction of those frustrated, including some with a history of mental illness and or substance abuse, will cope by aggressing against persons in their environment. The

4 persons who suffer or observe this aggression report it to authorities and the incidence of coerced treatment presumably increases.
We argue that the theory also implies that frustration induced by a contracting economy will lead some in the community to aggress against noisome and behaviorally impaired persons they earlier tolerated. Indeed, one of the earliest and most widely cited tests of frustration aggression theory implies such a mechanism. In their seminal paper on lynching of African Americans in the United States from 1882 to 1930, Hovland and Sears (1940) argued that many persons in a community feel frustrated when economic contraction halts or reverses improvement in income. The authors theorized that some whites in the
South would manifest such frustration by aggressing against African Americans believed to have threatened the white community. Hovland and Sears reported an inverse relationship between the price of cotton and the incidence of lynching.
Researchers, applying increasingly sophisticated methods to the lynching data, have replicated the original findings albeit with attenuated effects (e.g., Green,
Glaser, and Rich 1998; Hepworth and West 1988; Mintz 1946).
Sociologists (e.g., Blalock 1967; Blumer 1958; Kinloch 1974) and others
(Vanneman and Pettigrew 1972) extended the Hovland and Sears argument to offer an intolerance mechanism in which frustration induced in the larger community by increased competition for income manifests itself with aggression aimed at those, especially minorities, viewed as hostile to the social order.
Minority persons with severe mental illness likely fall into this category (Garland et al. 1998; Lindsey and Paul 1989; Rosenfeld 1984; Snowden 1999).

5
Overt physical or verbal aggression against noisome persons is not an option for law abiding persons, or for those who fear confrontation. These persons, however, can aggress in a relatively safe and societally acceptable way by calling the attention of authorities to noisome and threatening persons. Such aggression may not only provide relief from frustration, but also offer the possibility that authorities will remove the offending person from the environment.
The frustration and aggression that leads to notification of authorities may be impersonal as when a merchant wants a disabled alcohol abuser removed from the vicinity of a business. Or, it can be more personal as when a neighbor or caregiver brings an impaired and noisome person to the attention of the authorities. A caregiver may characterize his or her decision to alert authorities as one born more of frustration or lack of resources than of aggression (Hoge et al. 1998). Regardless of how one characterizes these decisions, however, they imply that those who made them no longer tolerate a previously acceptable circumstance.
A long-standing sociological literature makes the theoretical and empirical argument that society often applies the label "mentally ill" to persons for reasons other than their behavior (e.g., Gove 1975). Rushing (1978), for example, reported that persons with relatively little income and education were more likely to be coerced into treatment than persons with similar deviance but greater resources. Rushing infers that society has relatively little tolerance for the deviance of lower status persons. Combining this work with that suggesting lower tolerance for deviance among minorities during periods of economic

6 contraction (e.g., Blalock 1967; Blumer 1958; Kinloch 1974), suggests that behaviorally impaired minorities will be at greatest risk of coerced treatment during times of economic contraction.
We contribute to the literature by offering an empirical assessment of the plausibility of the reduced tolerance argument. More specifically, we test two hypotheses. The first is that the number of coerced admissions in the State of
California will vary inversely with the number of unemployed persons. We test this hypothesis because there would be little justification for testing the reduced tolerance argument if the phenomenon it seeks to explain does not replicate beyond San Francisco and Pittsburg (Catalano, Novaco, and McConnell 1997;
2002).
Our second, and principal, hypothesis is that the median level of functioning among African Americans coerced into treatment varies positively with unemployment in California. As described below, clinicians in California rate all persons receiving publicly financed treatment on a scale that measures how well the patient functions at intake regardless of diagnostic category or other circumstances that lead to treatment. We assume an essentially normal distribution of functioning in the community. Persons at the low end of the distribution will most likely experience coerced treatment because they appear most impaired. If the level of functioning at which the community presented candidates to clinicians were constant, the median functioning score among those coerced would remain constant even if a contracting economy pushed an increasing numbers of persons beneath that level. If the intolerance mechanism

7 were at work, however, a contracting economy would move the criterion for coercion upward (i.e., better functioning persons would be at elevated risk of coerced treatment). The median level of functioning of those coerced would therefore rise.
We test our hypotheses among African American and non-Hispanic White men and women over the age of 17. We separated the test population by gender because, as noted above, earlier research reported stronger effects for men
(Catalano, Novaco, and McConnell 1997; 2002).
METHODS
Data
Table 1 shows the monthly means for each of our test series. Data describing the functioning of persons coerced into treatment determined the length of our time series. The data series began July 1985 and ended June 1998
(i.e., the last month of data available to us at the time of the analyses).
Table 1 About Here
We used the monthly median of the Global Assessment Scale (GAS) and
Global Assessment of Functioning (GAF) scores as our measure of functioning in each of the monthly cohorts of persons coerced into treatment. The California
State Department of Mental Health requires clinicians at agencies providing publicly financed psychiatric services to record these scores upon intake. The

8
GAS measures overall psychosocial functioning, including psychological symptoms, as well as social and occupational functioning (Endicott et al. 1976).
The GAF, a slightly modified version of the GAS, was incorporated into the
American Psychiatric Association’s “Diagnostic and Statistical Manual (DSM-III-
R) in 1987 as Axis V, a formal component of diagnosis (American Psychiatric
Association 1987; 1994).
GAS and GAF scores range from 1 to 100. The lowest scores (i.e., 1-10) imply “persistent danger of severely hurting self or others or persistent inability to maintain minimal personal hygiene” while the highest (i.e., 90-100) imply
“superior functioning in a wide range of activities.” Psychiatric inpatients typically have scores between 1 and 40 (Spitzer et al. 1994).
The GAF is widely used in both research and clinical practice as a global measure of functioning. Both the GAS and the GAF have demonstrated adequate reliability and validity (Endicott et al. 1976; Goldman, Skodol and Lave
1992; Patterson and Lee 1995; Hilsenroth et al. 2000). Scores typically correlate well with other measures of clinical status (Roy-Byrne et al. 1996; Hilsenroth et al. 2000).
We used the number of unemployed persons in California to measure the economic climate. A person must not be working but looking for work to be counted as unemployed. The United States Bureau of Labor Statistics (2001) uses results from the Current Population Survey to estimate the number of unemployed persons in a state.

9
Our use of unemployment as an independent variable does not mean we assume that only unemployed persons and their families become frustrated due to economic contraction. We use unemployment because much literature (e.g.,
Dehejia and Rowe 1998) suggests that the macro economic circumstances that increase unemployment also impede upward mobility, measured as increased income and improved occupational status, among many who stay employed and their families.
Design
Data are consistent with causation when the dependent variable differs, as predicted by theory, from the values expected under the null hypothesis (i.e., under the assumption that the independent variable has no effect). The true experiment derives expected values under the null hypothesis from a randomly selected control group. We cannot randomly assign individuals to control and test communities in which we manipulate the macro economy. Tests of the effect of economic contraction on tolerance for deviance among minorities therefore must use statistical control. Such tests assume that the expected value of the dependent variable under the null hypothesis is the mean of its observations.
Time series, however, often exhibit trends, cycles, and the tendency to remain elevated or depressed after high or low values. These patterns, referred to collectively as autocorrelation, complicate tests because the expected value of autocorrelated series is not their mean (Catalano, Dooley, and Jackson 1983).

10
Sociologists typically respond to the problem of autocorrelation by identifying patterns and expressing them as an effect of earlier values, or "lags," of the dependent variable itself (Catalano, Hansen, and Hartig 1999; Catalano and Hartig 2001; Catalano and McConnell 1999; McCleary and Hay 1980). The residuals from a time-series equation with the correct lags exhibit no autocorrelation. The analyst can therefore add the independent variables to the equation to determine if their coefficients differ from zero in the hypothesized direction.
Any approach that removes autocorrelation from the dependent variable before testing the effect of the independent variable has the added benefit of avoiding type I errors due to shared trends and cycles (Catalano and Serxner
1987). The estimated coefficient for the independent variable is net of autocorrelation shared by the dependent and independent variables.
The strategy described above requires estimating the following equation.
 d
Y t
 ω
B n  d
X t



1


1
θB q
φB p





 a t
 d
is the difference operator that indicates a series was differenced at lag d (i.e., values at month t are subtracted from those at month t-d) to remove secular trends or seasonal cycles (i.e., render the series stationary in its mean).

11
Yt is, depending on the test, either the incidence of coerced treatment or the median GAF score in California during month t for one of the test groups.

is the effect coefficient for Xt.
Bn is the "backshift operator" that indicates a variable has been lagged n months. For the
 coefficient, Bn implies the association between y at month t and X at month t-n.
Xt is the number of unemployed persons in month t.

is the "moving average" parameter. Moving average parameters express the number of months that perturbations typically last.

is the autoregressive parameter. Autoregressive parameters gauge the number of months required for a series to revert to pre perturbation levels after a perturbation ends.
Bp and Bq are the backshift operators for the error term and indicate the lags at which the autoregressive and moving average effects begin. at is the error term at month t.
Analyses
We used the augmented Dickey-Fuller test to determine if equation 1 needed the  n
operators (Dickey and Fuller 1979). We used the strategy attributed to Box and Jenkins (Box, Jenkins, and Reinsel 1994) to identify and model autocorrelation (as expressed by the B,

, and
 terms in equation 1) in

12 the dependent variables. The strategy, Autoregressive, Integrated, Moving
Average (i.e., ARIMA) modeling, allows the researcher to empirically fit any of a large family of possible models to a time series. ARIMA models can be thought of as mathematical expressions of various filters through which temporally unpatterned series pass (Catalano et al . 1983). Each filter imposes a unique pattern of autocorrelation. In the Box-Jenkins approach, the researcher uses an iterative model-building process to infer the filter that imposed the autocorrelation observed in a time series.
We specified Equation 1 for coerced admissions by adding the X term to the equation resulting from the above steps. We included 3 "lags" (i.e., the n in

Bn was set to 0, 1, 2, and 3) to ensure that we allowed enough time for rising unemployment to induce awareness of increased competition, subsequent frustration, and, finally, coping through substance abuse or aggression. We then deleted insignificant parameters and estimated the remaining effects.
We followed the same general procedure for tests in which the median
GAF scores were the dependent variable. The intolerance argument, however, implies that unemployment should predict GAF scores at the same lag it predicts the incidence of coerced treatment. Our tests of the unemployment to GAF association, therefore, included only those lags, if any, found to be significant
(i.e., p < .05, two-tailed test) in the equations predicting coerced admissions.
RESULTS

13
Table 2 shows the estimated coefficients for equations predicting coerced admissions. As indicated by the ARIMA parameters, admissions for each group exhibited autocorrelation. Coerced admissions for all groups trended upward requiring that the difference operator be included for each series. The monthly differences for each group also exhibited "memory" in that large or small differences preceded elevated or depressed differences that became smaller in absolute size with passing months. The memory lasted one month for all groups while non-Hispanic whites exhibited additional autocorrelation in which large or small values predicted attenuated but similar movements a year later. There are many plausible explanations for this autocorrelation, but these are only incidental to the objective of this paper. Our objective requires that we control autocorrelation to avoid type I errors induced by shared cycles and other repeating patterns.
As shown in table 2, the incidence of coerced treatment for all groups rose above or fell below expected values 2 months after unemployment exhibited similar deviations. The number of unemployed persons increased in 58 of the
156 months studied. The average increase in these 58 months was 68,580.
This implies that labor market contraction was associated with about 1,435 coerced admissions of African American (i.e., .3609 X 68.58 X 58 = 1,435) and about 3,676 coerced admissions of non-Hispanic white (i.e., .9243 X 68.58 X 58
= 3,676) men during the 13-year test period. The estimates are 762 and 4,347 respectively for African American and non-Hispanic white women.

14
We conducted two additional analyses to assess the likelihood of a type I error. First, we estimated the final equations using natural logarithms of the coerced treatment variables to determine if temporal variation in variability could have affected the results. The results of the tests, with the obvious exception of the metric of the logged variables, did not change.
We also used the method devised by Hillmer (1984) and Alwan and
Roberts (1985) to determine if outliers had distorted the estimation of parameters. The method detected and controlled several outliers but the results of the test did not change.
Unemployment predicted the functioning scores for African American males only. Table 3 shows the coefficients for the final equation. As hypothesized, the median level of functioning among African American males mimicked earlier movements in unemployment at the same lag as did coerced treatment (i.e., two months).
We used the methods alluded to above (i.e., Alwan and Roberts 1985;
Hillmer 1984) to determine if outliers had distorted the estimation of parameters for GAF scores. The replacement of the GAS with the GAF, for example, could have induced slight level shifts in the time series. The method detected and controlled several outliers but the results of the test did not change.
The results in Table 3 support the hypothesis that tolerance for deviance among African American males declines after increases in unemployment. They do not, however, gauge how much of the association between the economy and coerced treatment might be due to intolerance. We estimated that quantity by

15 adding the GAF series to the equation used to test hypothesis 1. The decline in the coefficient for unemployment approximates the fraction of the association between unemployment and coerced treatment that might be due to reduced tolerance. Adding the GAF score of African American males causes the coefficient for unemployment to decline by approximately 20% (i.e., from .3609 to
.2895) although the point estimate remains significant (p <.01, two tailed test).
DISCUSSION
The above findings suggest that the ecological association between unemployment and coerced treatment of African American males for mental illness may be due, in part, to intolerance induced by economic contraction. We acknowledge that these findings describe California for the period analyzed.
Only replication elsewhere can establish the external validity of our findings. We, however, know of no reason to believe that California's population or procedures for coerced treatment (Ross, Rothbard, and Schinnar 1996) differ from those in other states in ways that could induce our findings.
Further research should clarify whose tolerance declines with contraction of the economy. Because the seminal (i.e., Hovland and Sears 1940) and more recent (Green et al. 1998) empirical work in the field focuses on aggression by majorities on minorities, intuition might suggest that intolerance among persons other than African Americans would account for our findings. Our theory, however, implies that a contracting economy could also affect the tolerance of
African Americans for mental illness in their communities and families. The low

16 level of assets available to African American households (United States Census
Bureau 1999) implies that they have few resources with which to assist a troubled family member or friend when economic contraction reduces income through lost hours of work or lost jobs. The loss may force African Americans to present a behaviorally impaired member of the household to mental health authorities in the hope of defraying the cost of maintaining him or her. Lost work hours or jobs may also lead to more time spent at home or in the community and therefore to greater contact with mentally ill persons. This added contact may increase dissatisfaction with, or awareness of, mentally ill persons and lead to their presentation to police or emergency medical personnel.
Hacker (1995) reported that African American women appear less threatening to the remainder of the community than do African American males.
Our findings suggest that this difference may extend to perceptions of danger to self and disablement.
Our findings may also help explain the patterns of treatment experienced by African Americans with mental illness. African Americans with mental illness disproportionately receive emergency treatment. The United States Surgeon
General (2001, p. 68) has noted the importance of this fact for improving the mental health of African Americans by writing: “Mental health care occurs relatively frequently in emergency rooms and psychiatric hospitals. These settings and patterns of treatment undermine delivery of high-quality mental health care.”

17
Attempts to explain African American psychiatric emergency room use typically focus on the notion that mentally ill African Americans lack access to outpatient care and therefore have a relatively high risk of becoming acutely ill.
Our findings suggest that societal intolerance may also play an important role.
Further research should determine which of the traditional categories of threats seems most affected by intolerance induced by economic contraction.
Intuition may suggest that danger to self would be least affected, but whether danger to others or gravely disabled should be more affected seems less obvious. The classic argument that frustration leads to aggression against those seen as threats suggests that persons threatening to others would be the obvious target of declining tolerance. One could argue, however, that relatively few persons threatening to others go undetected implying that the pool of such persons available for increased coercion may be small. A relatively large fraction of those who meet the gravely disabled criterion, however, may be mere nuisances under ordinary circumstances. The pool of them available for increased levels of coercion, therefore, may be relatively large at times of declining tolerance. A slowing economy may depress, for example, the tolerance of businesses for mentally ill persons who repel customers.
Further research should also focus on whether the associations we discovered have counterparts in the criminal justice system. Liska et al. (1999), reported that rates of psychiatric admissions rose with rates of jail admissions in a sample of American cities. Both correlated with jail capacity and, especially in segregated cities, the percentage of African Americans in the population. As a

18 city’s African American population increased, so did its jail capacity and, by way of increased jail capacity, admissions were higher both to jails and psychiatric facilities. In cities with greater African American representation, increased jail capacity facilitated use of the jail as a holding facility for transfer to psychiatric hospitals for psychiatric treatment. We hope our findings stimulate attempts to measure the association between economic contraction and the flow of persons between incarceration and mental health treatment.

19
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Table 1. Means and standard deviations (in parentheses) for times series used the tests (n = 156 months).
Coerced Admissions
Men
Non-Hispanic White
African American
Women
Series
Non-Hispanic White
African American
Median Functioning Score
Men
Non-Hispanic White
African American
Women
Non-Hispanic White
African American
Unemployed Persons (in 1000's)
Mean
3385 (453)
1027 (189)
2844 (375)
829 (132)
33.23 (2.13)
30.09 (1.10)
33.52 (2.19)
29.69 (1.77)
1053 (249)
28

Table 2. Final model coefficients for monthly incidence of coerced admissions in California for 156 months beginning July 1985 (Standard errors in parentheses).
Autocorrelation
Moving Average
Earlier
Lag 1
Autoregression
Lag 12
Unemployed Persons
(1,000's) 2 Months
*p<.05; two-tailed test
Men
Non-Hispanic white
Women
.4913* (.0762)
.2428* (.0870)
.9243* (.3553)
.4495* (.0797)
.3595* (.0837)
1.0930* (.3203)
Men
African American
Women
.5910* (.0671) .6306* (.0634)
.3609* (.1027) .1918* (.0818)
29

Table 3. Final model of median functioning score for African American males coerced in treatment in California for 156 months beginning July 1985 (Standard errors in parentheses).
Constant
Autocorrelation
Autoregression at Lag 1
Unemployed Persons (1,000's) 2 Months Earlier
*p<.05; two-tailed test
30.0981* (.1497)
.4643* (.0718)
.0022* (.0011)
30