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. Author manuscript; available in PMC: 2012 Mar 1.
Published in final edited form as: Fam Process. 2011 Mar;50(1):77–91. doi: 10.1111/j.1545-5300.2010.01347.x

Family Structure and Family Processes in Mexican American Families

Katharine H Zeiders 1, Mark W Roosa 1, Jenn-Yun Tein 1
PMCID: PMC3077030  NIHMSID: NIHMS255435  PMID: 21361925

Abstract

Despite increases in single-parent families among Mexican Americans (MA), few studies have examined the association of family structure and family adjustment. Utilizing a diverse sample of 738 Mexican American families (21.7% single parent), the current study examined differences across family structure on early adolescent outcomes, family functioning, and parent-child relationship variables. Results revealed that early adolescents in single parent families reported greater school misconduct, CD/ODD and MDD symptoms, and greater parent-child conflict than their counterparts in two parent families. Single parent mothers reported greater economic hardship, depression and family stress. Family stress and parent-child conflict emerged as significant mediators of the association between family structure and early adolescent outcomes, suggesting important processes linking MA single parent families and adolescent adjustment.

Youth living in single parent families are at greater risk than children in two parent families for academic (e.g., McLanahan, et al., 1994) and conduct problems (Brown, 2001), substance abuse, and depression (e.g., Compas & Williams, 1990). These well documented differences between single and two-parent families likely are related to challenges, opportunities, and interactions associated with family structure (See Barber & Demo, 2006 for a review). Thus, some studies have begun identifying the potential processes that help explain such differences. To date, most research has been conducted among European Americans or African Americans with little attention given to the Latino population, the largest minority group in the United States. This oversight is surprising given the significant increase in single-headed Latino households over the last two decades (Landale, Oropesa, & Bradatan, 2006). Among Mexican Americans specifically, single-parent families increased from 14% in 1980 to 22% in 2001. Of the few studies that have examined family structure among Latinos or Mexican Americans, inconsistent findings have emerged. Some studies suggest that Latino youth in single parent families are at greater risk for negative outcomes (e.g., Creighton, Park, & Teruel, 2009; Gil, Vega, Biafora, 1999), while others suggest they are not (e.g., Battle, 2002; Foster & Kalil, 2007). Inconsistencies could be partially due to the fact that many studies examining Latino families, and Mexican American families specifically, have not utilized samples that vary by income, language preference, nativity and even cultural values. An examination of more representative samples of Mexican American families is greatly needed to understand the impact of family structure on important familial processes.

Most theoretical explanations for the effects of family structure focus on family context processes that differ between single and two parent families (See Amato, 1993 for review). For instance, the parental loss perspective posits that children in single parent families exhibit more negative outcomes than children in two parent families because family members lack relational and financial resources that are common in two parent families. In contrast, the economic hardship perspective argues that negative outcomes in single parent families are due primarily to reduced economic resources (Amato, 1993). Finally, the family conflict hypothesis argues that both interparental and parent-child conflict can help explain differences by family structure.

Consistent with these theoretical ideas, several studies suggest that there are important mediators of the relation between family structure and outcomes. For instance, single parent families often report lower income and greater financial strain than two parent families (e.g., Avison, Ali, & Walters, 2007) and these factors help to explain differences in adolescent outcomes by family structure (e.g., Carlson & Corcoran, 2001). Further, maternal depression and mothers’ reports of their own stressors within the family are greater in single parent families and also help explain the differences in adolescent outcomes (Barret & Turner, 2006). Research has also shown important differences in parent-child interactions or processes between single and two parent families. For instance, Baer (1999) found that adolescents in single parent families reported more conflict, less positive communication, and lower levels of family cohesion than in two parent families. Although Dunifon and colleagues (2002) found that these differences in parenting variables did not mediate the association between family structure and adolescent outcomes, some findings have suggested that they do (Demo & Acock 1996). However, because of the cultural importance of the family in the lives of Mexican Americans (Marín & Marín, 1991), the effects of family structure and the processes involved might not be the same (Gonzales, Knight, Morgan-Lopez, Saenz, & Sirolli, 2002). Some scholars posit that single parenthood might be less detrimental to Latino families because of the strong reliance on extended family members (Heard, 2007), while others have suggested that the effects might be stronger given that single parenthood violates the traditional two parent family norms (Gamble & Dalla, 1998). Surprisingly, very few studies have examined possible differences in Mexican American mothers or adolescents related to family structure or the family processes that might help explain differences. The current study addresses this limitation by examining multiple mediators of the relation between family structure and early adolescent outcomes. It is important to note that most of the research on mediators related to family structure have included only one or a few theorized mediators and/or tested them separately, not considering the role of each mediator in the context or presence of other meditational processes. Testing of numerous mediators simultaneously is a more realistic test of the real world context for families (numerous processes happening simultaneously) and is likely to advance our theoretical understanding of family structure effects.

While examining potential mediators of family structure effects, we cannot overlook that processes might differ by adolescent gender. For instance, males are thought to have a more difficult time adjusting in single mother families than females because they lack a same-sex parent who acts as a role model showing them appropriate gendered behavior (Demo & Acock, 1988). While evidence among other ethnic groups has been somewhat inconsistent (e.g. Dunifon et al., 2002; VanderValk, Spruijt, DeGoede, Meeus, & Maas, 2004), the examination of gender among Mexican American youth appears especially important given the cultural emphasis on traditional gender roles (Gowan & Trevio, 1998). Latino parents with strong gender role attitudes are more likely to engage in gender socialization with the same sex adolescent (Raffaelli & Ontai, 2004). Thus, Mexican American males in mother-only single parent families might be at a disadvantage because they lack a male figure.

Current Study

The current study takes a step towards understanding family structure processes in a more representative sample of Mexican American families than used in previous studies by examining the following questions: 1) Do Mexican American single parent families differ from two-parent families on youth outcomes [i.e., school misconduct, conduct disorder/oppositional deviant disorder (CD/ODD) symptoms, major depressive disorder (MDD) symptoms, and academic performance], family functioning (i.e., economic hardship, maternal depression, family stress) and parent-child process variables (i.e., monitoring, parent-child conflict, parent-child relationship quality)? 2) Do family functioning and parent-child relationship variables mediate the relation between family structure and youth outcomes? 3) Do these family processes differ by adolescent gender?

Method

The data came from a study of the role of culture and context in the lives of Mexican American families in a large southwestern metropolitan area (Authors citation, 2008). Participants were 750 Mexican American students in 5th grade and their families who were selected from schools that served very diverse communities. To be eligible (a) families had to have a fifth grader attending a sampled school; (b) both mother and child had to agree to participate; (c) the mother was the child’s biological mother, lived with the child, and self-identified as Mexican or Mexican American; (d) the child’s biological father was of Mexican origin; (e) the child was not severely learning disabled; and (e) no step-father or mother’s boyfriend was living with the child. One side effect of the eligibility criteria was that all cohabiting couples in the sample consisted of the child’s biological parents.

Twelve families were excluded from analyses because mothers reported their marital status as widowed. Previous literature suggests that that widowhood is a different phenomenon than other types of single-parent families (Biblarz & Gottainer, 2000). Among participating families, 160 (21.7%) were single-parent (mother only) families and 578 (78.3%) were two-parent families. Mothers who identified their marital status as (1) Never married and not living with partner (n= 59), (2) Married but not living together (n= 40) and (3) Divorced (n= 61), were treated as single-parents. If mothers reported (4) Living with a partner but not legally married (n= 79) or (5) Married and living together (n=499), they were considered to be in two-parent families. Some scholars have argued for the importance of differentiating cohabiting families from married families (Brown, 2004; Dunifon & Kowaleski-Jones, 2002). However, the current study included both types of families in the two parent category because the unmarried male partners are the biological parent of the child. Further, cohabitation or consensual unions are common, carry a different meaning, and often function as surrogate marriages in Latin cultures more so than they do in US mainstream culture (Castro Martin, 2002).

Family incomes ranged from less than $5,000 to more than $95,000 (average of $30,000 – $35,000). The mean age of mothers was 35.8 (SD = 5.77) and they averaged 10.3 (SD = 3.68) years of education. The mean youth age was 10.4 (SD = .55). Nearly 70% of mothers were interviewed in Spanish, while 82% of youth were interviewed in English. A majority of mothers (74.3%), but only 29.7% of youth were born in Mexico.

Procedures

The complete research procedures are described elsewhere (Author Citation, 2008). Here we summarize key features of these procedures. The original research team identified communities served by 47 public, religious, and charter schools from throughout the metropolitan area chosen to represent the area’s economic, cultural, and social diversity. Recruitment materials were sent home with all 5th grade children in these schools. Recruitment materials explained the project and asked parents to provide contact information if interested in participating in the study. Over 85% of those who returned contact information were eligible for screening (e.g., Latino) and 1,028 met eligibility criteria. Computer Assisted Personal Interviews lasting about 2 ½ hours were completed with 750 families, 73% of those eligible. These interviews were conducted by trained interviewers; each interviewer received 40 hours of training which included information on project goals and characteristics of the target population. Question and response option were read aloud in the participants’ preferred language. Participants were paid $45 each. For $10, teachers completed a questionnaire reporting on children’s classroom behavior and academic performance.

Measures

Adolescent school misconduct

For early adolescents’ classroom behaviors, we examined teachers’ reports of acting out behavior using a 6-item subscale of the Teacher-Child Rating Scale (T-CRS; Hightower et al., 1986) which includes items such as “Disturbs others while working” and “Overly aggressive to peers.” Teachers respond using a Likert scale ranging from (1) not a problem to (5) very serious problem. The T-CRS has been used in a Latino sample and demonstrated good internal consistency (.85; Spomer & Cowen, 2001). For the current study, Cronbach’s alpha was .92 for the T-CRS subscale used.

Adolescent CD/ODD and MDD symptoms

We used symptoms counts from the computerized version of the Diagnostic Interview Schedule for Children (DISC-IV;Shaffer, Fisher, Lucas, Dulcan, & Schwab-Stone, 2000) to assess CD, ODD, and MDD symptoms. Given that CD and ODD often co-occur and ODD is thought of as a precursor to CD, these symptom counts were summed into a combined CD/ODD score. Consistent with previous use of the DISC, a mean of youth and mother’s report of symptoms were used with higher scores reflecting greater CD/ODD and MDD symptoms (Shaffer et al., 1996). Fathers did not complete the DISC.

Adolescent academic performance rating

Teachers ranked youth’s academic performance in comparison with the others in the classroom. Responses ranged from (1) far below average/the bottom 1/5 of the class to (5) excellent/the top 1/5 of the class.

Economic hardship

Three scales were used to examine economic hardship: Inability to make ends meet (2 items), Not enough money for necessities (7 items), and Financial strain (2 items; Conger & Elder, 1994). These scales operate equivalently across ethnicities (European American vs. Mexican American) and language use (English vs. Spanish; Barrera, Caples, & Tein, 2001). The mean of the three scales was computed for mothers and fathers in two parent families, with higher scores representing greater economic hardship. Mothers’ scores alone were used for single-parent families. Cronbach’s alpha for combined mother and father reports of economic hardship was .86 and for mother report was .78.

Maternal depression

The 20-item Center for Epidemiologic Studies Depression Scale (Radloff, 1977) was used to assess maternal depression in the past month. This scale has been used extensively in the Latino population, with equivalence tests suggesting that it operates similarly in European Americans and Mexican Americans (Crockett, Randall, Shen, Russell, & Driscoll, 2005). Mothers responded to statements such as, “You were bothered by things that usually don’t bother you”. Responses ranged from (0) rarely or none of the time to (3) most or all of the time. Cronbach’s alpha was .91.

Family stress

To assess family stress, 8 items from the family trouble/change subscale of the Multicultural Event Scale for Adolescents (MESA; Gonzales, Tein, Sandler, & Friedman, 2001) were revised. The MESA was originally developed a largely minority urban sample that included a substantial number of Mexican American adolescents to assess commonly occurring stressors. For the current study, however, mothers were asked the same items (e.g., “Someone in your family had a serious medical problem or mental illness”). Mothers indicated whether each event happened within the last 3 months, with scores reflecting the count of events endorsed.

Parental monitoring

Adolescents responded to a 10-item scale assessing their mothers’ and fathers’ monitoring behaviors (e.g., “Your mother/father knows what you were doing after school”; Dumka, Gonzales, Bonds & Millsap, 2009). Responses ranged from (1) Almost never or never to (5) Almost always or always. For adolescents in two parent homes, we created a combined score of adolescents’ reports of both mothers and fathers monitoring because they were highly correlated (r = .76, p < .001). For adolescents in single parent families, adolescent’s reports of mother’s monitoring were used. This scale has exhibited good internal consistency in previous studies examining Latinos (Dumka et al.) and in the current study (mother’s α =.80; father’s α = .76).

Parent-adolescent conflict

To assess parent-adolescent conflict, we utilized the 10-item Frequency Assessment subscale from the Parent-Adolescent Conflict Scale (PACS; Ruiz & Gonzales, 1998). Items for the PAC were generated from qualitative research with Anglo, African American, Mexican American Spanish-speaking, and Mexican American English-speaking families with children. Using 10-items, adolescents reported on the frequency of conflict with their parent(s) (e.g., “You and your mother/father yelled or raised your voices at each other”) using a Likert scale ranging from (1) almost never or never to (5) almost always or always. In two parent families, adolescents’ reports of conflict with mothers and fathers were combined (r = .68, p < .001). For adolescents in single parent families, adolescents’ reports of conflict with mothers were used. The scale demonstrated acceptable reliabilities (mother’s α =.73; father’s α = .72).

Parent-adolescent relationship quality

Parent-adolescent relationship quality was assessed by asking the adolescent the following question (Matthews, Wickerama, & Conger, 1996): What kind of relationship do you have with your mother/father? Adolescents responded using a Likert scale ranging from (1) the worst to (7) the best. In two parent families, relationship quality with mothers and fathers (r = .67, p < .001) were combined to represent overall relationship quality with parents. In single parent families, relationship quality with mothers was used.

Plan of Analysis

To examine mean level differences by family structure, we conducted a series of Analysis of Variance (ANOVA) tests. To examine the processes that could explain family structure differences on adolescent outcomes, we utilized structural equation modeling (SEM) using MPLUS Version 5.1 (Muthén & Muthén, 2007). Multiple fit indices (CFI, RMSEA, and SRMR) were used to evaluate fit. Good (acceptable) model fit is reflected by a CFI greater than .95 (.90), RMSEA less than .05 (.08), and SRMR less than .05 (.08; Hu & Bentler, 1999). Mediation effects were tested using the product of coefficients method with the multivariate delta method of deriving the standard error (Sobel, 1982). The moderating role of adolescent gender was examined using multi-group analyses; a non-constrained model was run (all parameters were allowed to vary across groups) followed by a restricted model in which all parameters were constrained to be equal across groups. A chi-square difference test was conducted; however, given that this test is sensitive to sample size and model complexities we followed recent recommendations by Chen (2007). This involves following up a significant chi-square difference tests by examining differences in CFI, RMSEA and SRMR across the restricted and full model; a significant chi-square difference test along with CFI changes less than or equal to −.005 and a change of greater than or equal to .010 in RMSEA or .025 in SRMR would indicate noninvariance across models. To account for missing data, all MPLUS analyses utilized full information maximum likelihood estimation (FIML; see Schafer & Graham, 2002).

Results

Mean Level Differences and Correlations

Because of unequal sample sizes between single and two parent families, we examined the assumption of homogeneity of variances in the ANOVAs and found that the assumption was violated as indicated by a significant Levene’s Test for all variables expect academic ranking, monitoring, and parent-child relationship quality. Although Lindman (1974) showed that the F statistic is quite robust against violation of the assumption, to be conservative, we used p ≤ .01 as the criterion for identifying significant differences on variables that violated this assumption (Allen, Titsworth, & Hunt, 2008). Table 1 presents the means of all variables and proportion of variance explained by family structure (η�) for all significant differences. Results suggested that adolescents in single parent families showed greater school misconduct [F (1,717) = 8.88, p < .01], CD/ODD symptoms [F (1,736) = 10.06, p < .01], and MDD symptoms [F (1,736) = 12.52, p < .001] than their counterparts in two parent families. Differences on academic ranking did not emerge. For family functioning variables, single parent mothers reported greater depression [F (1,735) = 19.28, p < .001], family stress [F (1,737) = 61.40, p < .001], and economic hardship [F (1,733) = 28.88, p < .001]. For the parent-child variables, adolescents in single parent families reported greater parent-child conflict than single parent families [F (1,737) = 6.10, p ≤ .01]; however, no differences emerged on monitoring and parent-child relationship quality. Table 2 presents bivariate correlations among study variables.

Table 1.

Means of study variables

Variable Single Parent Mean Two Parent Mean
Adolescent outcomes
  †School misconduct (teacher report)** η2 = .012 1.77 1.54
  †CD/ODD symptoms (parent and child report)** η2 = .013 2.24 1.68
  Academic ranking (teacher report) 3.04 3.26
†MDD symptoms (parent and child report)*** η2 = .017 3.83 2.99
Family functioning variables
  †Economic hardship (mother report)*** η2 = .038 2.71 2.35
  †Maternal depression (mother report)*** η2 = .026 1.93 1.73
  †Family stress (mother report)*** η2 = .077 2.19 1.25
Parent-child variables
  Monitoring (child report) 4.22 4.17
  †PC conflict (child report)** η2 = .008 2.04 1.93
  PC relationship quality (child report) 6.44 6.41
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Note.

†

signifies that the variable violated the homogeneity of variance assumption, tested using Levene’s Test. These variables are only considered significant if p ≤ .01.

**

means differ at p ≤ .01,

***

p < .001. Across ANOVAs, sample size for single parent families ranged from 156 – 160 families and 562 – 578 for two parent families.

Table 2.

Correlations among study variables (N = 710)

Measure 1 2 3 4 5 6 7 8 9 10
1. Family Structure
2. Economic hardship   .19**
3. Maternal depression   .15**   .46**
4. Family stress   .29**   .09*   .19**
5. Monitoring   .03 −.08* −.06 −.01
6. PC conflict   .10** −.03   .03   .33** −.01
7. PC relationship quality   .02 −.04 −.02 −.01   .33** −.19**
8. Adolescent school misconduct   .11**   .08*   .09*   .10* −.01   .06 −.04
9. Adolescent CD/ODD symptoms   .12** −.01   .10*   .15* −.11**   .25 −.21**   .28**
10. Adolescent academic rating −.08* −.19** −.16** −.12** −.11** −.01   .04 −.39** −.06
11. Adolescent MDD symptoms   .14**   .08*   .13**   .06 −.08*   .21** −.09*   .16*   .41** −.09*
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Note:

**

p < .01,

*

p < .05. Family structure (1 = single parent, 0 = two parent)

Test of Process Model

Given the interest in understanding how mediators help to explain difference by family structure on adolescent outcomes, the next set of analyses only included outcomes in which differences by family structure were observed (i.e., school misconduct, CD/ODD symptoms, MDD symptoms). All family functioning and parent-child variables were included in the analyses even though differences did no emerge on all variables (i.e., monitoring, parent-child relationship quality) because they acted as controls (e.g., controlling for levels of monitoring and parent child relationship quality). Figure 1 presents the tested model. Consistent with our ANOVA analysis and because of unequal sample size, paths were considered significant only if the p value was equal to or less than .01. The fit of the hypothesized model was good [χ2 (9) = 18.60, p < .05; CFI = 0.986; RMSEA = 0.038; SRMR = 0.021]. As seen in Figure 2, single parent families (1= single parent, 0 = two parent) reported greater economic hardship, maternal depression, family stress, and parent child conflict. In the presences of (or controlling for) other mediators, economic hardship and maternal depression were not related to outcome variables. Family stress, however, predicted greater adolescent CD/ODD symptoms. Parent child conflict predicted greater adolescent CD/ODD symptoms and MDD symptoms, while parent child relationship quality predicted lower levels of symptoms. Mediational testing revealed that family stress mediated the relation between single parent families and CD/ODD symptoms [95% confidence interval (CI) = .03, .14]. Parent child conflict mediated the relation between family structure and CD/ODD symptoms (95% CI = .01, .05) and the relation between single parent families and MDD symptoms (95% CI = .01, .08). It is worth noting that the only direct relation still significant after accounting for mediation was the relation between family structure and MDD symptoms [c′ (direct effect) = .23, p = .01]. The model was examined to see if paths differed by adolescent gender. A significant chi-square difference test emerged [χ2 Δ (27) = 41.52, p < .05] and the change in CFI was −.021. However, changes in either the RMSEA (.001) or the SRMR (.017) did not meet the recommended values of change (.010, .025, respectively; Chen, 2007). Therefore, we concluded that the model did not differ by adolescent gender.

Figure 1.

Figure 1

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Hypothesized process model to explain family structure differences.

Figure 2.

Figure 2

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Results of process model examining mediators of the relation between family structure and adolescent outcomes (N = 738).

Note. ** p < .01, *** p < .001. Family functioning variables and parent-child variables were allowed to correlate in the analysis: economic hardship and maternal depression (.43, p < .001); maternal depression and family stress (.13, p < .001); family stress and economic hardship (.03, p = .234); PC conflict and monitoring (−.01, p = .750), monitoring and PC relationship quality (.34, p < .001); PC relationship quality and conflict (−.18, p < .001). Correlations among endogenous variables are as follows: School misconduct and CD/ODD symptoms (.24, p < .001); school misconduct and MDD symptoms (.12, p < .01); MDD symptoms and CD/ODD symptoms (.32, p < .001).

Discussion

Despite the growing population of single parent families among Latinos, little research has examined family structure effects and processes in Latino families, and virtually no research has been conducted utilizing a representative sample of Mexican Americans. Our study addressed this gap by focusing on differences on adjustment between single and two parent Mexican American families, and potential mediators of these differences. Our results revealed that Mexican American single parent and two parent families differed on outcomes in a manner consistent with previous findings among European American and African American families (e.g., Avison et al., 2007). Adolescents in single parent Mexican American families reported greater school misconduct, CD/ODD symptoms, and MDD symptoms than their counterparts in two parent families, but did not differ on academic rating. Mexican American single parent families reported greater economic hardship, maternal depression, family stress, and conflict than their two parent counterparts, but no differences were observed on parental monitoring or parent-child relationship quality. These differences (and lack of differences) highlight that single-parent Mexican American families face difficulties like other single parents, but key parent-child factors (monitoring, relationship quality) are important and less influenced by family structure. This could partially be due to the strong sense of family (or familism) exhibited in Latino cultures (Fuligni & Pederson, 2002) which have been found to be a protective factor for developing youth (e.g., Germán, Gonzales, & Dumka, 2009; Rodriquez, Mira, Paez, & Myers, 2007). That is, despite the considerable challenges faced by Mexican American mothers in single-parent families, their commitment to their family and their sense of responsibility for their family may reduce the effect of these challenges on their early adolescent children. It also is possible that the strong sense of collectivism in Mexican American extended family networks and in communities dominated by Mexican Americans means that these single mothers receive support and resources (e.g., child care, emotional support, assistance in child monitoring) that compensate for the challenges these mothers experience . Such an explanation is speculative; future studies should consider examining how Mexican American cultural values (a) differ across single and two parent families and (b) modify family processes in two parent and single parent families.

In an attempt to better understand the role of familial process in single and two parent families, we examined family functioning variables and parent-child process variables as possible mediators between family structure and outcomes. Results revealed that family stress and parent-child conflict mediated the relation between family structure and CD/ODD symptoms, while parent-child conflict emerged as an important mediator for MDD symptoms. Consistent with prior research among European Americans (e.g., Barber & Demo, 2006; Demo & Acock, 1996) our findings suggest that there are important family processes helping to explain family structure differences in Mexican American families. Parent-child conflict appears especially important, relating to increases in both CD/ODD and MDD symptoms. The family conflict hypothesis posits that conflict within the family is an important predictor of adolescent well being and can help explain differences seen across single and two parent families. In line with this, Mexican American adolescents living in single parent families reported greater conflict with their parent and in turn were at greater risk than their counterparts in two parent families. In contrast, little support emerged for the economic hardship perspective; single mothers did report greater hardship, but it did not emerge as a significant mediator. Family stress emerged as the only other significant mediator, helping to explain differences in adolescents’ CD/ODD symptoms.

In addition to providing support that important family processes mediate the relation between family structure and adolescent outcomes, our results also suggest some specificity in mediating processes. That is, certain mediators or family processes differentially influence specific adolescent outcomes. It is worth noting that some mediating variables were related to the adolescent outcomes in the zero-order correlations (Table 2; e.g., economic hardship & school misconduct), but did not emerge as mediators supporting the importance of testing multiple mediators simultaneously. Given that all the mediators were considered in one model, the test was a more stringent and realistic test of the mediating process since these processes happen simultaneously in a larger context. Therefore, our results suggest that in the context of a variety of mediators, family stress and parent-child relationship quality (which did not mediate, but was related to CD/ODD symptoms) were important for externalizing symptoms, while parent-child conflict related to both externalizing and internalizing symptoms.

Finally, in line with others (e.g., Demo & Acock, 1996) our results also suggest that family structure by itself lacks explanatory power for differences between Mexican American single and two parent families. The overall variance accounted for in adolescent outcomes by family structure was less than 2%. Rather, mediators, or family processes provided us with a better understanding of strengths and challenges these families face. Such findings underscore the importance of future studies to continue examining processes and in particular, cultural value processes in Mexican American families that might be either protecting families or placing them at risk for poor adjustment.

The current study represented an important step in identifying family structure differences and processes among Mexican American families by using a larger and more representative sample of Mexican Americans than had previously been studied and by simultaneously examining of a variety of potential mediating processes. In addition to research implications, our findings provide important information to clinicians and preventionist working with Mexican American families. These families face many challenges (economic, maternal mental health, family stress); however, counteracting such challenges are strong familial relationships between the adolescent and parent. With this in mind, clinicians can utilize a strength-based approach, focusing more on Mexican American families’ competencies, and unique cultural characteristics instead of deficits and challenges. Such an approach has emerged as a successful strategy in clinical work focused on decreasing youth emotional and behavioral problems (Cox. 2006). For prevention efforts, the current study identifies risk factors for Mexican American single parent families, some of which are malleable and potential targets for prevention programs. For instance, focusing on preventing parent-child conflict within Mexican American single-parent families could help decrease internalizing and externalizing symptoms. Finally, it is possible that both intervention and prevention efforts might want to consider tapping into the resources of these mothers’ extended family and community networks to facilitate their ability to meet their children’s needs.

Despite strengths, the current study it is not without limitations. First, the study relied upon cross-sectional data, limiting our conclusions about the direction of effects. Without longitudinal data, it is unclear if differences on outcomes can be attributed to the effects of family structure or if maternal and child problems contributed to the family’s current status. Future longitudinal research, ideally starting before family structure changes occur (in the case of divorce or separation), is needed to fully understand how family structure impacts youth development. Additionally, even though mediators were chosen based on theory, an examination across time provides a stronger test of meditational processes. Another limitation of the current study is that we lacked information about other adults outside of the immediate family that might influential in that adolescent’s life, either as a source of support or a source of conflict for adolescents (e.g., Barret & Turner, 2006). Future research should examine relationships both residential parent and additional adults (e.g., relatives) living in the home and how these relationships change overtime. Despite these limitations, the study provides an important first step in understanding family processes in single and two parent Mexican American families.

Acknowledgments

Work on this paper was supported, in part, by grant RO1 MH 068920 (Culture, context, and Mexican American mental health, Mark Roosa, Principal Investigator), the Cowden Fellowship program of the School of Social and Family Dynamics at Arizona State University, and a Training Grant T32MH18387. The content is solely the responsibility of the authors and does not necessarily represent the official views of the National Institute of Mental Health or the National Institutes of Health. The authors are thankful for the support of Marisela Torres, Jaimee Virgo, the La Familia Community Advisory Board, consultants, and interviewers, and the families and teachers who generously participated in the study.

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