Journal of Gerontology: SOCIAL SCIENCES 1992. Vol. 47, No. 2. S55-65

Copyright 1992 by The Gerontological Society of America

Perceived Intergenerational Solidarity and Psychological Distress Among Older Mexican Americans Renee H. Lawrence,1 Joan M. Bennett,2 and Kyriakos S. Markides3

A model separating and relating dimensions of intergenerational solidarity with measures of psychological distress was investigated for older Mexican Americans. Solidarity consisted of measures of similarity, affection, and association. Measures of psychological distress were somatic/retarded symptoms, depressed affect, and positive affect. To evaluate whether emotional closeness with a particular child modified the linkages, the model was analyzed separately based on whether or not the elderly participant reported that the child included in the intergenerational study was her or his closest child. The findings indicated that the impact of affection and association was a function of the particular dimension of distress and the emotional closeness of the child. Although the proposed model needs expanding, it provides some support for the expectation that family solidarity has important consequences for elderly Mexican Americans.

but growing number of studies have examined A SMALL the causal linkage between intergenerational solidarity and psychological well-being or distress among elderly

Markides and Mindel, 1987), is that, given the importance of familial contact and solidarity, the greater the perceived solidarity with family members the less the psychological

Mexican Americans. However, limited attention has fo-

distress experienced by elderly persons (Sussman, 1985).

cused on theoretical models linking intergenerational solidarity and psychological distress and on factors that may modify the linkages (cf., Bengtson and Kuypers, 1985; Roberts and Bengtson, 1990). Accordingly, this investigation addresses two specific conceptual problems. Thefirstis the lack of a well-articulated model depicting the linkages among dimensions of intergenerational solidarity and dimensions of psychological distress. The second is a failure to ascertain the impact of emotional closeness with a child on these linkages for older Mexican Americans. The concept of intergenerational solidarity has been used to describe the nature of the special bond among family members (Bengtson etal., 1985; Brubaker, 1990; Sussman, 1985) with recent attention turning toward ethnicity as an important consideration in intergenerational research (e.g., Mindel, Habenstein, and Wright, 1989; Stanford, Peddecord, and Lockery, 1990). In spite of Mexican Americans' increased urbanization, this group continues to have a more cohesive family support system than other groups, with faceto-face contact between family members remaining especially important (Becerra, 1988; Markides and Mindel, 1987). Such social support is viewed as having important positive effects on emotional well-being, as interaction with important others leads to a sense of self-worth and value (Bengtson, Olander, and Haddad, 1976; Cohen and Wills, 1985; Krause, Liang, and Yatomi, 1989; Pearlin et al., 1981). The expectation, especially regarding Mexican Americans (Markides, Costley, and Rodriguez, 1981;

Despite the expectation that more solidarity is associated with a psychological advantage, the evidence to date is equivocal in general (cf., Bengtson and Kuypers, 1985) and in particular for Mexican Americans. Markides, Costley, and Rodriguez (1981) found that perceived solidarity with adult children had a significant independent and positive effect on the morale of elderly Mexican Americans. In another study, Markides and Krause (1985) found that association and affection with adult children had no significant effect on life satisfaction for elderly Mexican Americans independent of other important factors. They also unexpectedly found no effect for affection toward the child on feelings of depression. Further, they found that greater association was related to more depression in elderly Mexican Americans even after attempts to statistically control for dependency on their children. The evidence strongly indicates a need to qualify simple notions about the impact of solidarity upon psychological well-being (Bengtson and Kuypers, 1985; Hagestad, 1984). One qualification concerns the simple assumption of a norm of equal attachment with respect to all children. Several researchers have begun to question this norm and suggest that parents' feelings of closeness or emotional attachment to their children vary (Aldous, Klaus, and Klein, 1985; Cicirelli, 1983). Parents do have preferences and are selective in their relationships with their various children (Aldous, 1987; Cicirelli, 1990). Thus, a legitimate consideration derived from attachment theory is the evaluation of how S55

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'Philadelphia Geriatric Center. institute of Gerontology, University of Michigan. 'Department of Preventive Medicine and Community Health, University of Texas Medical Branch, Galveston.

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emotional closeness with a particular child influences the effects of intergenerational solidarity (e.g., visiting, dining together) on psychological distress.

AFFECTION

SOMATIC SYMPTOMS

SIMILARITY DEPRESSED AFFECT

POSITIVE AFFECT ASSOCIATION Figure 1. Proposed model depicting the linkages among components of intergenerational solidarity and psychological distress.

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Modeling Intergenerational Solidarity and Psychological Distress There are many aspects or dimensions of intergenerational solidarity, and viewing them as separate but interrelated components is one way of examining the effects of solidarity on psychological distress (e.g., Markides and Krause, 1985). Recent theory has focused on identifying aspects or dimensions of intergenerational cohesion (e.g., Mangen, Bengtson, and Landry, 1988). Bengtson and Schrader (1982) and McChesney and Bengtson (1988) discuss several constructs related to intergenerational solidarity, three of which have been identified as core aspects for indexing cohesion: affection, association, and attitudinal consensus or similarity (see also Roberts and Bengtson, 1990). Although there is recognition of the multidimensionality of solidarity, little has been done to establish a model interrelating the various dimensions of solidarity (Atkinson, Kivett, and Campbell, 1986; McChesney and Bengtson, 1988; Roberts and Bengtson, 1990). For example, when studying Mexican Americans, Markides, Costley, and Rodriguez (1981) aggregated dimensions of solidarity (association and affection), whereas Markides and Krause (1985) separated the dimensions but did not assess the complexity and interrelationships among the multiple dimensions. Accordingly, rather than representing solidarity as a simple composite, the present study proposes a basic causal model (Figure 1) disaggregating the three core dimensions' of solidarity (cf., Atkinson, Kivett, and Campbell, 1986; Roberts and Bengtson, 1990). Specifically, the model represents similarity with a particular child (perceived degree of similar values and attitudes) as causally linked to both affection (nature and extent of positive sentiment among family members) and association (extent to which family members share activities with other family members) with that child. However, association and affection are not related

in a causal fashion but are correlated. The linkages are based on the theoretical assumptions that the more perceived similarity, the more association and affection (Heider, 1958). Affection and association were not causally linked because some have suggested that association induces affection (e.g., Heider, 1958;Homans, 1950) while others (including Heider) have suggested that affection induces association (Roberts and Bengtson, 1990). Psychological distress is also clearly recognized as a multidimensional construct although there is little consensus on the number and nature of the dimensions (cf., George and Bearon, 1980; Kane and Kane, 1981). With regard to the effects of intergenerational solidarity on psychological distress, inconsistency and lack of clarity in findings may in part result from inattention to the unique effects of intergenerational solidarity measures on different dimensions of psychological distress. Some researchers have noted the advantages of distinguishing dimensions of psychological distress (e.g., depressed affect vs somatic symptoms) rather than relying on a composite scale score (cf., Berry, Storandt, and Coyne, 1984; Blazer, 1981; Newmann, 1984). Accordingly, causally separating different aspects of familial solidarity with respect to the different dimensions of psychological distress may clarify and isolate effects. For example, the finding by Markides and Krause (1985) that association with children was related to greater distress may hold for some but not all dimensions of distress. Therefore, the model (Figure 1) depicts three aspects of psychological distress as differentially linked to affection and association. Specifically, somatic/retarded symptoms (physiological manifestations of psychological distress), depressed affect (feelings of sadness and depression), and positive affect (lack of feelings of happiness and hopefulness) are represented as independent but interrelated dimensions of distress (cf., Liang et al., 1989). Moreover, similarity is hypothesized to have no direct impact on psychological distress, only indirect via the other measures of solidarity (affection and association). As mentioned earlier, the other major purpose is to deter-

INTERGENERATIONAL SOLIDARITY

METHODS

Participants The data came from a three-generation study of Mexican Americans in San Antonio, Texas, collected during 1981-82 (Markides et al., 1983). Area probability sampling involved selection of census tracts and city blocks within census tracts to identify Mexican Americans aged 65 to 80 who had an adult child and grandchild (18 years old or over) living in the metropolitan area. When two or more such three-generation lineages were available, one was selected randomly. Of the 375 older participants, 321 were included in this study. Reasons for exclusion from the present analyses were having only one child (n = 27) (i.e., no choice regarding whether the child is the closest child or not), lack of information about whether a child was designated as the closest child (n = 14), and/or missing data on more than two relevant variables (n = 13). The controlled pairwise deletion strategy for missing data resulted in a range for the total sample of 295-321. Subsamples for the group analyses were determined through response to the question, "Of all of your children, with whom do you have the closest relationship?" (child not in the study, child in the study, or equally close to all your children). There were 117 (pairwise range: 107-127) older Mexican Americans in this sample for whom the closest child was not in the study ("Not Closest" group); 68 (pairwise range: 62-73) identified the target child as the closest child ("Closest" group); and 111 (pairwise range: 101-121) reported being equally close to all their children ("Equally Close" group). A critical issue in intergenerational research deals with living arrangements. When evaluating residency for this sample of Mexican Americans, only 20 of the respondents

were sharing residence with the adult child included in the study. A significant chi-square (9.66, df = 2, p = .008) indicated that there was a relationship between living arrangement (shared residence with the adult child vs not) and group (Closest, Not Closest, and Equally Close). Specifically, the chi-square indicated that more of the subjects than expected by chance lived with the closest child (n = 9 or 12.3%), and significantly fewer than chance lived with the not closest child (n = 2 or 1.6%). For elderly persons claiming to be equally close to all children, the percent of shared residency was about what is expected by chance (n = 9 or 7.4%). In order to evaluate the impact of shared residency, analysis of the model (as reported in Results) was done deleting those 20 respondents, and the findings were basically comparable to the full sample results. Accordingly, only the findings including all elderly respondents are reported below. Markides et al. (1983) discussed the issue of sample representativeness in three-generation studies. Three-generation studies have been criticized for containing a disproportionate number of older-generation persons who are more traditional (e.g., married younger), more fertile, and overrepresent women (Bytheway, 1977), all of which are true of this sample (see Table 1). In addition, these elderly Mexican American three-generation family members had significantly fewer years of education than elderly Mexican Americans not having three-generation families (Markides et al.). Although it is important to keep these sample characteristics and considerations in mind while interpreting the results, it is also true that the target population was defined as threegeneration families containing elderly Mexican Americans (cf., Hill, 1977). Measures Survey items for multi-item scales were evaluated by confirmatory factor analysis, and only those items demonstrating equivalence across the three groups of elderly respondents (Not Closest, Closest, and Equally Close) were retained in the current analysis (Joreskog, 1971; Liang et al., 1989). Without establishing such equivalence, comparisons of the causal linkages would be ambiguous and misleading (Baltes and Nesselroade, 1970). Moreover, because of the constraints of the sample sizes it is necessary to use composites in analyzing the model. Use of composites (Liang et al., 1990) reduces the number of parameters that are estimated to only those linkages represented in Figure 1. Reliability estimates (total coefficient of determination, Joreskog and Sorbom, 1989) for the measures are reported based on only those items equivalent across these three groups of elderly respondents. Intergenerational solidarity. — Three dimensions of intergenerational solidarity were used: Affection, Association, and Similarity. All the measures used are based on the reports provided by the older Mexican Americans and thus represent their perceptions of the relationship with the adult child included in the study. The items used were derived from the Southern California Study of Generations (Bengtson and Schrader, 1982). Affection assessed the degree of sentiment or "subjective

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mine whether emotional closeness with the child is relevant to understanding the effects of solidarity on psychological distress (cf., Aldous, Klaus, and Klein, 1985; Quinn, 1983). If closeness is important, not directly addressing it may confound results and our understanding of the effects of solidarity on well-being. That is, the linkages between measures of intergenerational solidarity and psychological distress may depend on whether or not the child is reported as the closest to the respondent from among her or his children. It seems reasonable to assume that the emotional outcomes or consequences of interacting with the closest child may be different than interacting with a child who is not the closest. Similarly, concerns are raised when the questions addressing solidarity ask for a summary measure across all of the elderly participant's adult children (e.g., Lee and Ellithorpe, 1982; Markides, Costley, and Rodriguez, 1981). The information is, in essence, the average for all children and, though providing one important type of information, it does not allow the assessment of subtle and potentially critical differences engendered by emotional closeness with a particular child. Accordingly, this research addresses solidarity regarding one particular child of the elderly respondent and whether or not that child is perceived as the closest of his or her children.

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Psychological distress. — Psychological distress was measured by the Center for Epidemiologic Studies Depression Scale (CES-D) (Radloff, 1977). The structure of psychological distress used is based on Liang et al. (1989), who reported a model that adequately fit the data for the older generation in this study. Responses represent how often the symptom was felt during the past week and were scored 0 (rarely or none of the time/less than 1 day) to 3 (most or all of the time/5-7 days). All items (including "positive affect" items) are coded such that higher scores indicate more psychological distress. The first dimension, SomaticlRetarded Symptoms, contains four items representing physiological manifestations of psychological distress (poor appetite, restless sleep, unable to "get going," and trouble concentrating). All items reflected the construct in a similar manner for the three subgroups. Reliability estimates were .618 for Not Closest, .732 for the Closest, and .734 for Equally Close. The second dimension, Depressed Affect, was assessed by five items including inability to shake off blues, crying spells, and feelings of depression, loneliness, and sadness. All items were equivalent across the subgroups except for "crying spells," and therefore the item was deleted. Reliabilities of the scale for the remaining four items were .838

for Not Closest, .871 for Closest, and .856 for Equally Close. The third factor, Positive Affect, contains three items assessing feelings of happiness, hopefulness, and enjoyment of life. No significant differences were found for these items. Reliabilities for this index were .753 for Not Closest, .824 for Closest, and .651 for Equally Close. Control variables. — In order to evaluate the effects of intergenerational solidarity on dimensions of psychological distress, other important correlates of distress were introduced as controls (Diener, 1984; Larson, 1978). The variables were age (in number of years), sex, marital status (currently not married vs married), education (actual number of years of school), total monthly income (seven categories ranging from less than $200 to greater than $1200), and health (rated as excellent, good, fair, or poor, ranging from 1 to 4 with a higher score reflecting poorer health). To further clarify the nature of the effects of solidarity on psychological distress, three instrumental dependency measures were also included in the process of evaluating the model. The items are the same as the three used by Markides and Krause (1985) to assess help received by the older generation from the child in the study: (1) amount of help with chores or errands (ranging from 1 "almost never or never" to 8 "almost every day"), (2) help when sick (1 "never when I am sick" to 4 "every time I am sick"), and (3) help with financial matters (1 "no, not at all" to 4 "regularly receive most of my support from him or her"). Analysis Strategy Two general data analytic steps were used in analyzing the linkages between family solidarity and psychological distress: (1) descriptive statistics and group difference analyses of the variables (Descriptive Analysis section, below); and (2) analysis of the model including examination of group differences in parameter estimates (Evaluation of the Proposed Model section, below). Specifically, the variables and model were evaluated separately for the three groups of older Mexican Americans: Closest, Not Closest, and Equally Close. The model depicted in Figure 1 was estimated using the structural equation program LISREL (Joreskog and Sorbom, 1989). In order to obtain the best estimates possible, reliability estimates were used to adjust for measurement error (see Liang et al., 1990) because ignoring measurement error leads to biased and inconsistent estimates and to inaccurate assessment of the relations between theoretical constructs (Bollen, 1989). In addition, residualized covariance matrices were used to remove the effects of the control variables (income, self-rated health, education, age, sex, and marital status). Throughout the process, covariance matrices were used for comparing the three groups to ensure that the comparisons were not confounded by variance differences between groups (Blalock, 1967). Measurement equivalence was also imposed to ensure that any differences found could not be attributed to differences across groups in the measurement properties of the constructs. It is important to recall that previous analyses were undertaken to ensure the feasibility

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interaction" between the participant and the adult child in the study. Six items asked how well the elderly respondent understood, respected, and trusted the adult child and how well the elderly respondent felt understood, respected, and trusted by that child. Responses were coded 1 (not well) to 6 (extremely well). Only the four items assessing respect and trust were found to be equivalent and therefore were included in the final scale. Reliability estimates were .873 for the Not Closest group, .820 for the Closest group, and .876 for the Equally Close group. Association items (seven) measured "objective" interactions and activities, including visits just to talk, recreation outside the home, family gatherings, discussions of matters important to the older respondent, attending religious activities together, telephone conversations, and dining together. Responses were coded 1 (almost never or never) to 8 (almost daily). Five items were found to be equivalent measures of the construct across the subgroups ("telephone conversations" and "attending religious activities together" had to be deleted). Reliability estimates for the groups were .818 for Not Closest, .720 for Closest, and .811 for Equally Close. Similarity was measured by one item assessing overall similarity of views about life between the elderly respondent and the adult child ("In general, how similar are your views about life to those of ?") using a scale ranging from 1 (very different) to 6 (extremely similar). Reliability for similarity (assessed by one item only) was constrained to 0.8. Thus, although it was not possible to estimate the reliability of this single-item construct, given the above information on the reliability of the measures of affection and association, 0.8 seemed to be a reasonable estimate in lieu of explicit information and was more realistic than the common practice of assuming the measure to be perfectly reliable.

INTERGENERATIONAL SOLIDARITY

RESULTS

Descriptive Analysis Table 1 summarizes the descriptive information for the variables. Separate multivariate analyses of variance (MANOVAs) were used to assess significant differences among the groups for the control, solidarity, psychological distress, and dependency variables as sets. The multivariate tests were nonsignificant for the solidarity measures (F = 1.45, df = 6,622, p = .189), the background control variables (F = 1.72, df = 8,584, p = .091), and the psychological distress scales (F = 1.25, df = 6,616, p = .279). However, multivariate tests were significant for the dependency control variables (F = 2.68, df = 6,512, p = .014). Univariate analyses (df = 2,259) revealed that "chores" (F = 4.86,/? = .008) and "illness" (F = 5.60,p = .004) differed significantly among groups, whereas "financial help" did not (F = .379, p = .685). Scheffe tests indicated that for both "chores" and "illness" the Closest group was helped by the child significantly more than the Not Closest group, with neither of these groups differing significantly from the Equally Close group. Differences among groups regarding composition of sex and marital status were assessed by chi-square analyses. Neither was significant (chi-square = .44, p = .80 for sex; chi-square = 3.54, p = .17 for marital status). Evaluation of the Proposed Model The proposed model (Figure 1), initially evaluated using the total sample (Table 2), exhibited a good fit (bottom, Table 2). The linkages among the solidarity measures were as expected: the more the perceived similarity, the more reported affection and association. Affection had a signifi-

cant and negative impact on both somatic/retarded symptoms and depressed affect, indicating that the more reported affection for the adult child, the less somatic symptoms and depressed affect. Association had a significant positive effect on the same two dimensions of distress: the more association, the more somatic/retarded symptoms and the more depressed affect. All of the correlated residuals among the dimensions of distress were significant, as was the residual between association and affection. For the multi-sample analyses, the proposed model exhibited a good fit for the groups when imposing equivalence constraints on the measurement aspects of the model. Table 2 (bottom) summarizes the various indices of overall fit. Although the measures of overall fit were good for the three groups, some differences did emerge. Table 2 presents the parameter estimates (unstandardized) for the causal linkages and correlated residuals. The linkages among the intergenerational solidarity measures were as expected although significant only for the Not Closest and the Equally Close groups. Specifically, the more perceived similarity between the respondent and the adult child, the more affection was felt and the more association reported. Moreover, affection and association were positively related. The model of solidarity hypothesizes that similarity will not have any significant direct effects on the dimensions of psychological distress. This was supported for all groups, including the total sample. Specifically, the residuals and the modification indices indicated that the direct effects are not significant and need not be incorporated in the model. The significant indirect effects of similarity were negative and related to somatic/retarded symptoms for the Not Closest and the Equally Close groups. The expected pattern of affection impacting negatively on psychological distress was found for the Not Closest and Equally Close groups. However, the only significant effect was found between affection and somatic/retarded symptoms: the more affection felt for the adult child, the fewer somatic/retarded symptoms reported. The pattern of causal linkages between association and the dimensions of psychological distress were predominantly positive. However, the only significant linkage was for the Not Closest group: the more association, the more somatic/ retarded symptoms reported. To identify nonequivalent causal linkages and nonequivalent correlated residuals, additional analyses were conducted. In particular, confidence intervals were estimated comparing each of the groups. The results can be summarized as follows: There were no statistically significant differences between the Not Closest and the Equally Close groups. However, the Closest group significantly differed from both of the other groups on the causal linkages from affection to somatic/retarded symptoms and the correlated residual between affection and association. In addition, the Closest group differed significantly from the Not Closest group on the linkage between somatic/retarded symptoms and positive affect. Differences in the linkage between similarity and affection were not quite significant. Thus, as can be seen in Table 2, for the total sample the magnitude of many linkages was reduced by inclusion of the

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of assuming that the measurement structure of the constructs was equivalent. The causal model was tested simultaneously for the three groups using the multi-sample analysis in LISREL (see Joreskog, 1971). For present purposes, this technique is preferable over assessing statistical interaction effects within an ordinary least squares regression framework for several reasons. First, the effects of measurement errors are compounded in cross-product terms; second, error in the component parts is correlated with error in the cross-product terms (Bohrnstedt and Marwell, 1978). Both of these considerations result in biased estimates (Busemeyer and Jones, 1983). Moreover, the present analysis strategy allows explicit testing of the proposed model by including the correlations between the measures of psychological distress and the correlation between association and affection (for further examples and discussions of the issues see Krause, Bennett, and van Tran, 1989, and Mutran and Reitzes, 1984). Several measures of the overall fit of the proposed models were applied, including the likelihood ratio and other measures of fit less affected by sample size. One such measure is the goodness-of-fit index (GFI) provided by Joreskog and Sorbom (1989). The Bentler-Bonett (1980) normed fit index (Delta,) and the Bollen non-normed fit index (Delta2) are two other measures used in the evaluation process (Bollen, 1989).

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Table 1. Descriptive Information

Variables (range) Intergenerational solidarity Similarity (1-6) M SD Affection (4-24) M SD

N* Sex

% Male % Female Marital status % Not married % Married Education (0-14) M Years SD Income (Monthly) M Category' SD N* Health rating (1-4) M SD Instrumental dependency Chores (1-8) M SD Illness (1-4) M SD Financial (1-4) M SD

Not Closest (N = 127)

Closest (N = 73)

Equally Close (N = 121)

3.33 1.19

3.22 1.28

3.52 1.16

3.31 1.10

19.81 2.77 317

19.39 2.92 125

20.34 2.35 73

19.92 2.79 119

22.20 6.50

21.09 6.61

23.48 5.39

22.60 6.83

1.99 2.25 319

2.00 2.11 125

2.11 2.39 73

1.90 2.33 121

2.04 2.65 320

2.03 2.56 127

2.58 3.01 73

1.72 2.46 120

2.21 2.20 317

2.31 2.20 124

2.31 2.52 72

2.06 2.01 121

74.34 4.76 319

74.41 4.89 126

74.21 4.87 73

74.34 4.58 120

27.7 72.3

26.0 74.0

27.4 72.6

29.8 70.2

51.1 48.9

53.5 46.5

57.5 42.5

44.6 55.4

2.78 3.07 316

2.96 3.00 126

3.31 3.54 71

2.26 2.79 119

2.56 1.10 307

2.59 1.20 121

2.61 1.21 70

2.49 .89 116

2.65 .87 320

2.67 .87 127

2.73 .85 73

2.58 .88 120

3.68 2.03 262

3.23 2.05 109

4.07 1.93 55

3.97 1.98 98

2.86 1.17

2.63 1.19

3.19 1.09

2.91 1.13

1.47

1.42 .60

1.53 .65

1.50 .61

.61

W's at top of the table represent upper value of the range for controlled pairwise deletion. N's under specific variables represent actual sample size for that variable when it was less than the upper value. ••Psychological distress items are scored so higher values indicate more distress. 'Income Categories: (1) $0-199; (2) $200-399; (3) $400-599; (4) $600-799; (5) $800-999; (6) $1000-1199; and (7) $1200+ .

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Association (6-48) M SD Psychological distress6 Somatic/retarded symptoms (0-12) M SD A"1 Depressed affect (0-12) M SD A"1 Positive affect (0-9) M SD N" Control Age (64-81) M Years SD

Total (N = 321)

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INTERGENERATIONAL SOLIDARITY

Table 2. Unstandardized Parameter Estimates (Maximum Likelihood) by Total Sample, Group, and Total Sample Without Closest Group Group Comparisons Linkage

P GFI Delta, Delta2

Not Closest

Excluding Closest Group

Closest

Equally Close

Rerun of Original

Controlling for Dependency

.452* .315*

.533* .390*

.242 .175

.506* .354*

.496* .337*

.409* .145

- .297* -.193* -.002

- .486* -.274 -.107

.184 -.002 .304

-.511* -.235 -.061

- .434* -.258* -.085

- .295* -.139 -.210

.233* .220* .021

.356* .279 .215

.403 .496 -.001

.227 .089 -.054

.469*

.425*

.169

.785* .241* .253*

.724* .016 .123

.863* .591* .527*

308.88 15 .000 .758

331.92 45 .000 .748

4.07 3 .254 .995 .987 .997

10.04 9 .347 .998 .970 .997

.715

.975

.268* .200 .062

.072 .093 .054

.452*

.493*

.449*

.851* .251 .184

.735* .139 .171*

.717* .320* .299*

.732

.988

246.28 15 .000 .750 1.63 3 .652 .998 .994 1.007

162.95 15 .000 .779 2.25

3 .522 .996 .986 1.005

*p < .05.

Closest group. This was particularly true on those linkages where pronounced differences emerged between the groups, indicating that important differences do exist and that the effects are interactive, not additive. Accordingly, no further analyses were done using the total sample collapsing across the three groups of closeness. Although it would have been ideal to keep the groups separate when evaluating the effects of the dependency measures, the reduction in sample sizes due to missing information on the dependency items precluded such a strategy (see Table 1). Therefore, in order to maximize sample sizes, additional analyses were undertaken by combining the Not Closest and the Equally Close groups and excluding the Closest group. This was done based on the above analyses, which indicated that the Not Closest and Equally Close groups exhibited similar patterns of significant effects in the model and lack of significant differences in the confidence interval analyses. Although additional analyses on the Closest group would be helpful, the reduction in sample size due to missing information on the dependency items and the differences between the Closest group and the other two groups precluded any further analysis of this group individually or in combination with the other groups. Prior to controlling for dependency, the original model was rerun so that clear comparisons could be made when introducing the dependency measures. The results for the

model excluding the Closest group are provided in Table 2 (N = 227, pairwise range: 224-231). The model exhibited a good fit for this combined analysis. Not surprisingly, the linkages were basically identical to those reported earlier for the two groups individually. In addition, affection's impact on depressed affect was now significant, and association had a positive significant effect on somatic/retarded symptoms (this was not previously true for both groups). The final analysis evaluated the model after residualizing the matrices for not only the effects of the original control variables but also for the effects of the three instrumental dependency variable (see Table 1 for list of the dependency measures). Residualized matrices were used instead of directly modeling the effects of the dependency variables, a strategy employed due to the sample sizes. Moreover, the concepts of dependency and interdependency are complex (Mancini and Blieszner, 1989) and it is not clear how best to model these effects. However, in family research, association has been found to be frequent, and with that association a variety of personal services are exchanged. The present goal was to understand the effects of dependency (upon the child in the study) on the linkages among solidarity and psychological distress. The last column in Table 2 presents these results. This analysis is based on a controlled pairwise sample size of 190 (range of 187-193) after introducing the dependency measures. The overall measures of fit indicate a good fit. Interestingly, the linkages from association to the

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Causal linkages Similarity to: Affection Association Affection to: Somatic/Retarded symptoms Depressed affect Positive affect Association to: Somatic/Retarded symptoms Depressed affect Positive affect Correlated Residuals Affection and association Somatic/Retarded symptoms and: Depressed affect Positive affect Depressed affect and positive affect Measures of Fit: Null Model Chi-square df P GFI Model in Figure 1 Chi-square df

Total Sample

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dimensions of psychological distress have been reduced to nearly nonexistent linkages, whereas the linkages from affection to the dimensions of psychological distress are negative. Although only one is statistically significant, there is an indication for the first time that an aspect of solidarity might be linked to positive affect. The correlated residuals among the dimensions of psychological distress are all significant as is the residual between affection and association. The relationship from similarity to association was substantially reduced and is no longer significant after controlling for instrumental dependency (two of which are types of association: "help with chores" and "illness"). DISCUSSION

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The study sought to clarify the discrepancy in findings regarding the causal linkages between perceived family solidarity and psychological distress among older Mexican Americans by proposing and evaluating a causal model separating and interrelating dimensions of solidarity and dimensions of distress. Another purpose was to evaluate how emotional closeness with the adult child modifies these causal linkages. The results from a reanalysis of data from a three-generation study used by Markides and Krause (1985) suggest that the proposed model is useful in identifying and isolating the effects of solidarity on psychological distress, and reinforce the need to attend to emotional closeness with the child as perceived by the elderly respondent. Specifically, the findings revealed that the expected negative effect of affection upon psychological distress was qualified by degree of closeness with the child and the particular dimensions of psychological distress. For respondents who reported about a Not Closest child or reported being Equally Close to all children, the more affection reported for that child, the less psychological distress. However, the only significant reduction in psychological distress was found for somatic/retarded symptoms. For the analysis combining these two groups, this significant effect (more affection related to fewer somatic/retarded symptoms) was reduced but remained even after controlling for dependency upon that adult child. The findings indicated that the impact of association was also a function of the particular dimension of distress and closeness with the child. The predominant pattern of effects suggests that the more the association, the greater the psychological distress, replicating Markides and Krause (1985). However, the present findings isolate that significant effect to the dimension of somatic/retarded symptoms for the Not Closest group, and for the combined group (i.e., Not Closest and Equally Close). Although the present model depicted association causally impacting on somatic/retarded symptoms, the opposite direction of impact needs to be mentioned: increased psychological distress, in particular as measured by somatic symptoms, may lead to increased association (cf., Mutran and Reitzes, 1984). It is most likely the case that the causal flows are in both directions, but longitudinal data are needed to evaluate this issue. In the analysis where the effects of instrumental dependency were removed, the linkage between association and somatic/retarded symptoms was no longer significant. Therefore, unlike Markides and Krause (1985), controlling

for instrumental dependency on the adult child in the survey reduced the effects of association with that child on the three dimensions of distress. In fact, the effects were reduced to near zero, suggesting that the effects observed for association for this sample of Mexican Americans are explained by instrumental dependency. That is, once measures of dependency were taken into account in the model, the significant linkages between association with the child and psychological distress were practically eliminated. In the present analysis, dependency items were introduced separately and only the items measuring dependency on the adult child were incorporated. The implication here is that creating a composite scale from the dependency items (e.g., Markides and Krause) may balance out effects and reduce the impact. Most of the significant effects of solidarity on distress involved the somatic/retarded symptoms dimension. In fact, after controlling for dependency, the only remaining significant effect was between affection and the somatic/retarded symptoms dimension. These findings are consistent with literature showing that older adults are more likely to express psychological distress in physiological terms than in positive affect or depressed affect terms (e.g., Berry, Storandt, and Coyne, 1984). This finding is also consistent with literature on the somatization of distress among Hispanics (e.g., Angel and Guarnaccia, 1989; Guarnaccia, Angel, and Worobey, 1989). That is, if elderly Mexican Americans tend to present psychosocial distress as a series of somatic complaints, then finding that more affection is related to fewer such complaints (after controlling for health ratings) reinforces the importance of distinguishing dimensions of distress and the buffering role of affection. None of the linkages among the solidarity measures and the dimensions of psychological distress was significant for the group of elderly respondents referencing the closest child. It should be noted that the Closest group had a small n and, accordingly, large standard errors. Thus, as seen in Table 2, although it appears that there are notable effects, the small n introduces some cautions regarding the stability of parameter estimates, despite the magnitudes of the estimates. Although sample size is of concern, it is still important to ask why causal relationships did not emerge. It is worth noting that the possible explanation of greater restriction in range for the solidarity measures for the Closest group is not a likely explanation for the lack of significance and/or differences (as can be seen in Table 1, the means and variances for the Closest group are not markedly different from the other two groups for the solidarity measures). The implication is that either there are no significant linkages or the proposed model is inadequate for representing solidarity with the closest child. Unfortunately, the survey used for the present analyses was not designed to evaluate differences in emotional closeness. Accordingly, it is not directly possible to investigate several issues that might explain the lack of findings for elderly persons reporting perceived similarity, affection, and association with the closest child. At least two related possibilities include: (a) attending to heterogeneity in the Closest group and (b) developing a more complex model. Regarding the issue of heterogeneity, the larger standard errors associated with the Closest group may be a function of

INTERGENERATIONAL SOLIDARITY

emonial interaction and informal interaction. There are at least two other areas where the model needs revising in order to explain the effects of intergenerational solidarity on psychological distress. One area deals with the absence of other critical variables in the model. For example, the proposed model focused only upon parent-child relations and did not incorporate any measures of nonfamilial relationships (cf., Quinn, 1983). Similarly, the present model did not depict relationships with other children or incorporate family structure variables, such as number of children. In addition, the linkages were modeled assuming solidarity causally impacts upon psychological distress. However, as mentioned earlier, it is as reasonable to assume that psychological distress impacts on aspects of solidarity. Such possibilities are confounded by reliance upon crosssectional data, and several researchers have noted the critical need for a longitudinal and developmental perspective (e.g., Lewis, 1990; Mancini and Blieszner, 1989). Another area focuses on the fact that the proposed model concentrated on the perceptions of the elderly participant and did not incorporate input from the adult child. Although considering only one individual is a reasonable starting point, it is far from an accurate understanding of a complex process. Variation in feelings of closeness and attachment between parents and children is a function of the behaviors and preferences of both parents and children (Hagestad, 1984). Nevertheless, it is important to obtain an understanding at the individual level because such information may affect the causal linkages and interpretations of dyadic-level processes (cf., Roberts and Bengtson, 1990). Future considerations should address congruence or lack of congruence between the parent and child on the measures of solidarity in order to capture dyadic-level processes. Despite the current limitations, this study makes several contributions. One arises from directly evaluating perceived closeness of the adult child. Parents do differentiate between their children, and this study indicates that, at least for Mexican Americans, there are different psychological effects associated with solidarity when attending to perceived closeness with a child. Moreover, the lack of findings for the effects of solidarity with a closest child indicates that more research is necessary to better understand this unique relationship. In addition, this study supplies further evidence for the theoretical multidimensionality of solidarity and the separate and differing effects of those dimensions (similarity, association, and affection) on various aspects of psychological distress. Although the proposed model clearly needs expanding and could benefit from direct comparisons with other ethnic groups, it nevertheless provided some support for the expectation that intergenerational solidarity has important consequences for this group of elderly Mexican Americans: for a child who is not the closest (Not Closest and Equally Close groups) the more affection reported for that child, the fewer somatic symptoms reported even after controlling for dependency on that child.

ACKNOWLEDGMENTS

This research was supported by the National Institute of Mental Health (grant number MH-44214) and the National Institute on Aging (grant

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meaningful variation rather than entirely due to small sample size considerations. That is, there may be less agreement among the respondents regarding what constitutes or defines a closest or favorite child, than with regard to identifying a child who is not the closest or claiming to be equally close to all children (e.g., Aldous, Klaus, and Klein, 1985). Indeed, the criterion for defining a child as favorite or closest may vary between elderly respondents as well as vary for the same respondent across time and/or situation (Hagestad, 1984). Accordingly, the pattern of causal linkages among the components of solidarity and the prediction of distress may be reduced due to differences in the meaning of a closest child. In contrast, for children who are clearly and consistently identified as not the closest, as similarity increases so does affection and association, with affection and association relating to some dimensions of distress. Moreover, it may be possible that when children and parents are extremely close, solidarity makes no additional contribution to explaining psychological distress or does not make a linear contribution. The Closest group may represent a smaller group of extremely close children and this may help explain why there were fewer closest children in this random sample of children. The data set did not permit direct assessment of these issues. It would be helpful to collect data that (a) assess degrees of closeness rather than the three general categories used here, (b) evaluate intrafamilial differences in closeness, and (c) assess differences in the definition of favorite or closest. Until a better understanding of closest is obtained, variations will appear as error variance in predictive models. Regarding the second possibility, it may be that the linkages between solidarity and psychological distress are more complex when discussing the closest child. The closeness of the relationship may add stresses that are not incorporated in the proposed simple model. Aldous (1987), not surprisingly, reported that as the frequency of troublesome emotional exchanges increased, the overall satisfaction with the relationship decreased. Such exchanges may be more frequent and/or stressful when they involve the closest child. What is needed is an analysis of the content of the interactions which contribute to the current measure of association. In general, further research and modeling efforts need to include a better understanding of the association measures. This is particularly true given the elimination of the significant linkages between association and psychological distress when introducing the dependency measures. In fact, the linkages between association and psychological distress were reduced to statistical nonexistence. The implication is that the association measures may not be clearly differentiated from functional solidarity items (helping/exchange of assistance and services). Improvement in understanding might be obtained by identifying and modeling additional dimensions of association beyond evaluating the content of the exchanges. For example, it might be worth attending to obligatory vs nonobligatory distinctions as well as selfinitiated vs other-initiated. One might reasonably wonder whether family gatherings (perhaps perceived as obligatory and other-initiated) are the same type of association as visits just to talk (which in some instances might be better represented as nonobligatory and self-initiated). In fact, Bengtson and Schrader (1982) suggested distinguishing between cer-

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number AG-09344). The original data collection was supported by the National Institute on Aging (grant AG-01573). The authors are grateful to Marcene Goodman, Sarah Haddad, Jennifer A. Klapper, Neal Krause, Powell Lawton, Jersey Liang, Linda Nyquist, and Paul Thuras for helpful comments. Address correspondence to Dr. Renee H. Lawrence, Behavioral Research, Philadelphia Geriatric Center, 5301 Old York Road, Philadelphia, PA 19141. REFERENCES

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Received September 13, 1990 Accepted August 15, 1991

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Mangen, Vern L. Bengtson, and Pierre H. Landry, Jr. (Eds.), Measurement of lntergenerational Relations. Beverly Hills, CA: Sage Publications. Mindel, Charles H., Robert W. Habenstein, and Roosevelt Wright, Jr. 1989. "Family Lifestyles of America's Ethnic Minorities: An Introduction." In Charles H. Mindel, Robert W. Habenstein, and Roosevelt Wright, Jr. (Eds.), Ethnic Families in America: Patterns and Variations (3rd ed.). New York: Elsevier. Mutran, Elizabeth and Donald C. Reitzes. 1984. "lntergenerational Support Activities and Well-being Among the Elderly: A Convergence of Exchange and Symbolic Interaction Perspectives." American Sociological Review 49:117-130. Newmann, Joy P. 1984. "Sex Differences in Symptoms of Depression: Clinical Disorder or Normal Distress?" Journal of Health and Social Behavior 25:136-159. Pearlin, Leonard I., Elizabeth G. Menaghan, Morton A. Lieberman, and Joseph T. Mullan. 1981. "The Stress Process." Journal of Health and Social Behavior 22:337-356. Quinn, William H. 1983. "Personal and Family Adjustment in Later Life." Journal of Marriage and the Family 45:57-73.

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