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Am J Public Health. 2008 January; 98(1): 125–132.
PMCID: PMC2156066

Maternal Depressive Symptoms, Parenting Self-Efficacy, and Child Growth


Objectives. We assessed whether maternal depressive symptoms and parenting self-efficacy were associated with child growth delay.

Methods. We collected data from a random sample of 595 low-income mothers and their children aged 6 to 24 months in Teresina, Piauí, Brazil, including information on sociodemographic characteristics, mothers’ depressive symptoms and parenting self-efficacy, and children’s anthropometric characteristics. We used adjusted logistic regression models in our analyses.

Results. Depressive symptoms among mothers were associated with 1.8 times higher odds (95% confidence interval [CI] = 1.1, 2.9) of short stature among children. Parenting self-efficacy was not associated with short stature, nor did it mediate or modify the relationship between depressive symptoms and short stature. Maternal depressive symptoms and self-efficacy were not related to child underweight.

Conclusions. Our results showed that among low-income Brazilian families maternal depressive symptoms, but not self-efficacy, were associated with short stature in children aged 6 to 24 months after adjustment for known predictors of growth.

Stunting caused by undernutrition disproportionately affects the poor,1 is indirectly responsible for more than half of deaths among young children worldwide,2 and is associated with inadequate physical and cognitive development and limited social mobility.36 Research from industrialized countries shows that failure to thrive (growth delay) is more likely to occur in children of depressed care-givers,6,7 but few studies have examined this relationship in developing countries.

Previous studies in which both short stature and underweight have been used as indicators suggest an association between maternal depression and inadequate growth.8 One investigation involving a clinic-based sample in Jamaica showed that rates of depression were higher and parenting self-esteem was lower in a group of mothers with underweight children than in a group of mothers with well-nourished children.9 In another study, psychiatric disorders among mothers living in favelas (shanty towns) in São Paulo were associated with child underweight.10 A study comparing 4 developing countries showed a relationship between maternal depression and child underweight in India and Vietnam but not in Peru or Ethiopia.11 Reasons for such regional variations are not well understood.

Psychosocial conditions have received increasing attention in efforts to explain children’s growth and development. The concept of collateral health effects (the transfer of health benefits from one individual to another through social rather than biological pathways12) may be pertinent because optimizing mothers’ mental health may foster positive caregiving behaviors that benefit their infants. Parenting self-efficacy (a parent’s level of confidence in performing in her or his role as a caregiver) and depression are closely related,1315 and high self-efficacy is related to a variety of positive parenting behaviors1619 (e.g., responsive feeding practices and breast-feeding duration) that may affect child growth.

Maternal psychosocial care has been shown to be associated with child nutritional status,20 but it may play a more important role in lower- and middle-income countries than in affluent countries.8 For example, maternal depressive symptoms and parenting self-efficacy may be especially salient in low-income populations facing economic stressors. We examined relationships between low-income Brazilian mothers’ depressive symptoms and self-efficacy as exposures and short stature and underweight among their infants as outcomes.


Study Site and Data Collection

Participants were 595 mothers of children aged 6 to 24 months. Mothers were randomly sampled from 9 low-income communities representing 4 geographic areas in Teresina, Piauí, Brazil, that were similar in terms of household income and neighborhood resources (see Surkan et al.21 for study details). Two of the 4 areas were chosen in part because they received services from the Programa de Saúde da Familia (Family Health Program), a government program that targets low-income families and provides local access to a team of health professionals, growth monitoring services, and health education.22 We identified 2 other areas that were scheduled to receive the program’s services in the near future (Maria Vieira, Fundação Municipal de Saúde de Teresina, oral communication, late 2001 or early 2002).

Data from the local sewage and water company were used to draw maps of the study areas using AutoCAD version 2000i (Autodesk Inc, San Rafael, Calif), and field workers updated maps as necessary. Initially, a census of approximately 8000 houses was conducted to identify 1432 eligible households (i.e., households with a child aged between 6 and 24 months and a primary caregiver older than 15 years). These households were assigned numbers between 1 and 1432 on the maps so that interviewers could locate them. Corresponding random numbers were used to sample approximately 150 of these houses from each of the 4 geographic areas. When there were 2 eligible children in a household, a coin toss determined the participating child.

Fifteen local female interviewers who had undergone training and a local study coordinator collected data in 2002. Training spanned 2 weeks, including practice with anthropometry and interviewing techniques. Selected households in which the caregiver was absent were approached 5 times on different days and different times of day before being removed from the study.

Children’s birth dates and birthweights were copied from a card (Cartão da Criança) on which medically relevant data are recorded by health personnel. Interviewers gathered information on breast-feeding duration, mother’s age and education, household possessions, living conditions, sanitation conditions, income, and participation in the Family Health Program, as well as other topics outside the scope of our study.

Dependent Variables

Short stature-for-age relative to reference growth curves is considered an indicator of chronic undernutrition caused by repeated infections, poor dietary quality, or both.3 Low weight relative to height, or wasting, is considered a growth indicator distinct from stunting, reflecting acute undernutrition or severe disease. Variance in weight reflects both stature and weight-for-height.3 Because the rate of wasting (0.9%; n = 5) in our sample was too low to allow use of this variable as an outcome, we assessed underweight after we controlled for stature-for-age.

A team of 2 interviewers used the Hanging Spring Dial Scale (Charles Morgan and Sons, London, England) to take weight measurements and the Lightweight Infantometer (Perspective Enterprises, Portage, Mich) to take length measurements (each interviewer was responsible for 1 type of measurement).23,24 Available measurements were averaged. The World Health Organization SAS (SAS Institute, Cary, NC) program based on Multicenter Growth Reference Study curves25 was used to calculate z scores for computing short stature and underweight (defined as less than −2 standard deviations of the reference height-forage and weight-for-age z scores, respectively). Anthropometric data were missing for 6 children.

Independent Variables

The primary independent variable was mothers’ score on the Center for Epidemiological Studies Depression Scale (CES-D),26 a 20-item scale (score range: 0–60) that assesses depressive symptoms over the previous week. We developed a Portuguese-language version of the CES-D (for details, see Surkan et al.21); scores were dichotomized, with scores of 16 or above corresponding to depressive symptomatology. The Cronbach alpha coefficient for our Portuguese CES-D was 0.82, comparable to the coefficient (0.88) for another version of the CES-D created in southeastern Brazil running concurrently with our study.27 Thirteen mothers had missing data on 1 of the scale items, and 1 mother had missing data on 2 items; for these women, the available items were averaged and scaled.

Parenting self-efficacy was assessed using a 10-item scale designed to measure mothers’ perceived beliefs regarding their performance in performing specific caregiving tasks with their baby.28 Women rated their own parenting ability on a 4-point scale (not at all good, not good enough, good enough, or very good). High and low categories of maternal self-efficacy were defined as above and below the median of the 10 summed items, respectively. Of the 595 participants, only 23 (4%) had missing data (on only 1 or 2 items). Thus, scores were based on the average of the items answered. The Cronbach alpha coefficient for the scale was 0.71, indicating satisfactory internal consistency among the items.

We used conventional cutoffs or response distributions to code demographic variables. The variables assessed were child gender, birthweight (less than 2500 g, 2500 g or more), and age (6–11 months, 12–17 months, 18–24 months); breast-feeding duration (less than 6 months, 6–11 months, 12 months or more); mother’s education (0–7 years, 8 years or more); and number of children living in the household (1, 2, or more). (Because only 7 women reported exclusive breast-feeding, the breast-feeding duration variable referred to any breast-feeding.)

We assessed families’ sanitation conditions with a scale composed of 5 items (using a water filter, having garbage collected, having a sewage system or toilet not connected to the sewage system but with water available, presence of a running water source in the house or yard, and owning a refrigerator). A score of 5 (reflecting positive responses to all 5 items) corresponded to good sanitation conditions, and a score of 0 corresponded to poor conditions. The scale was analyzed as a continuous variable. If any of the items were not answered (which occurred in 13 cases), the sanitation scale score was coded as missing.

We also created a composite measure that focused on socioeconomic status (SES) and living conditions. The variables assessed on this “SES and living conditions” scale included household income (0 = R $0–89, 1 = R $90–179, 2 = R $180–359, 3 = R $360 or more; the exchange rate at the time was approximately R $2.5 to US $1, and the minimum wage was R $180 per month); presence in the home of electricity, a fan, a radio, and a television (0 = presence of 0–2, 1 = presence of 3, 2 = presence of all 4), types of walls in home (0 = unfinished mud or plastic, 1 = finished mud, 2 = brick), type of flooring in home (0 = mud, 1 = cement or a combination of cement and mud, 2 = ceramic, cement, or a combination of cement and ceramic), roof type (0 = thatched, paper, or plastic, 1 = brick or concrete). Possible scores ranged from 0 (worst SES and living conditions) to 10 (best SES and living conditions). The SES and living conditions scale was modeled as a continuous variable.

Statistical Analysis

We used SAS version 9.13 (SAS Institute Inc, Cary, NC) to model the growth outcomes, short stature and underweight, as dichotomous variables. We conducted bivariate analyses examining the relationship of each independent variable with both outcomes. We calculated the χ2 test of association between growth outcomes and dichotomous categorical predictors. Trend tests were used to examine the relationships between SES and living conditions, and sanitation scale scores and both growth outcomes.

Covariates included in the multivariate models were child gender, birthweight, and age; breast-feeding duration; maternal educational attainment; sanitation scale score; SES and living conditions scale score; number of children living in the household; and participation in the Family Health Program. We first constructed a model with only these covariates and with short stature and underweight as outcomes. The CES-D and maternal self-efficacy measures were added into these models separately to test whether they were related to child growth after adjustment for covariates. We used PROC GENMOD in SAS (with interviewer included as a random effect) to create logistic regression models.29

To examine the mediating effect of maternal self-efficacy on the relationship between depressive symptoms and growth outcomes, we fit a model including maternal self-efficacy, depressive symptoms, and covariates. As a means of testing for potential effect modification, stratified models of short stature by maternal self-efficacy and CES-D score were fit. Because underweight was not related to either depressive symptoms or self-efficacy as a main effect, models assessing effect modification were fit only for short stature.


Of the 732 randomly selected families we attempted to enroll in the study, 613 agreed to participate; our final sample was limited to the 595 households in which mothers were the primary caregivers. Of the 119 families selected that did not participate, 67 had moved, we were not able to locate 21, 13 were not at home after 5 repeated visits, 8 were out of town for a prolonged period, and 6 refused to take part; also among these families, 2 children had died, 1 child was hospitalized, and 1 mother appeared unable to concentrate on the survey. The prevalence of short stature was 25%, and the prevalence of underweight was 4%. Fifty-six percent of mothers scored in the high depressive symptom range (16 or above) on the CES-D, whereas 49% fell into the category of low maternal self-efficacy.

Short Stature

In bivariate analyses of short stature (Table 1 [triangle]), 48% of low-birthweight children (less than 2500 g) were stunted, as compared with 22% of normal-birthweight children. Almost one third of children of mothers with less than 8 years of education were stunted. Slightly more than one third of children in households with the least optimal sanitation conditions and the lowest scores on the SES and living conditions scale were stunted. Among children of mothers with depressive symptoms, 30% were stunted, in comparison with only 18% of children of nondepressed mothers. The prevalence of short stature was the same among children of mothers with low and high self-efficacy scores (25%).

Relationships Between Demographic and Maternal Psychosocial Variables and Child Short Stature and Underweight: Teresina, Piauí, Brazil, 2002

Multivariate analyses (Table 2 [triangle]) showed that low maternal educational level, low SES and living conditions scale score, and low child birthweight were related to short stature. Children from households not participating in the Family Health Program had 0.7 lower odds of short stature relative to children from households participating in the program (95% confidence interval [CI] = 0.5, 1.0). When CES-D score was added to the multivariate model, high maternal depressive symptomatology was related to a nearly 2-times greater odds of short stature (odds ratio [OR] = 1.8; 95% CI = 1.1, 2.9) compared with children of mothers with low depressive symptomatology. Low maternal self-efficacy was not related to short stature (OR = 1.0; 95% CI = 0.8, 1.2).

Odds Ratios for Short Stature Among Children Aged 6 to 24 Months, by Sociodemographic Characteristics, Maternal Depressive Symptoms, and Maternal Self-Efficacy: Teresina, Piauí, Brazil, 2002

To assess whether self-efficacy modified the effects of depressive symptoms, we simultaneously added CES-D score and maternal self-efficacy to the model shown in Table 2 [triangle]. The addition of these variables did not result in any changes in the effect estimates for depressive symptoms (OR = 1.8; 95% CI = 1.1, 2.9) or self-efficacy (OR = 1.0; 95% CI = 0.8, 1.2).

Finally, we stratified the analyses by restricting the data to maternal self-efficacy scores above or below the median (data not shown). We found no change in the association of short stature with maternal self-efficacy according to whether mothers were classified as having (OR=1.0; 95% CI=0.6, 1.5) or not having (OR=0.8; 95% CI=0.4, 1.5) depressive symptoms. The effect of depressive symptoms on short stature was not dependent on maternal self-efficacy (OR=2.1; 95% CI=0.9, 4.9, for high self-efficacy and OR=1.7; 95% CI=1.1, 2.8, for low self-efficacy). Also, a test assessing whether there was an interaction between depressive symptoms and maternal self-efficacy showed no significant interaction effects on short stature (P=.9).


Prevalence of underweight (Table 1 [triangle]) increased from 3% among children aged 6 to 11 months to 5% among children aged 12 to 17 months. Only 2% of children breast-fed between the ages of 6 and 11 months were undernourished, as compared with 5% of children breast-fed for less than 6 months or for a year or more. Five percent of children whose mothers had completed less than 8 years of education were underweight, in comparison with 1% of children whose mothers had completed at least 8 years of education. Low scores on the SES and living conditions scale were associated with a greater prevalence of underweight. Six percent of children whose mothers had high CES-D scores were undernourished, as compared with 2% of children whose mothers did not exhibit depressive symptoms. The percentage of children who were underweight did not vary according to maternal self-efficacy.

In multivariate models including sociodemographic variables (Table 3 [triangle]), low birthweight was associated with more than 5-times greater odds of underweight (OR=5.5; 95% CI=2.4, 12.5), and better sanitation conditions were associated with reduced odds of underweight (OR = 0.6; 95% CI = 0.4, 0.8). Other covariates were not strongly or significantly related to underweight. When added to the adjusted model, neither CES-D score (OR = 1.8; 95% CI = 0.6, 5.5) nor maternal self-efficacy (OR = 1.3; 95% CI = 0.4, 4.5) was significantly associated with underweight (Table 3 [triangle]).

Odds Ratios for Underweight Among Children Aged 6 to 24 Months, by Sociodemographic Characteristics, Maternal Depressive Symptoms, and Maternal Self-Efficacy: Teresina, Piauí, Brazil, 2002


Our results suggest that, in low-income Brazilian families, compared with women who had low levels of depressive symptoms, high levels of maternal depressive symptoms were associated with approximately a 2-times greater risk of short stature in their children aged 6 to 24 months after adjustment for known determinants of growth. We found no association between maternal depressive symptoms or self-efficacy and underweight, nor did maternal self-efficacy appear to mediate or modify the relationship between depressive symptoms and short stature. In addition, the cross-sectional design of our study largely precludes us from making causal inferences.

Fifty-six percent of mothers displayed depressive symptoms at or above the traditional CES-D score cutoff of 16. Perhaps this high prevalence is not entirely surprising given that our population included mainly low-income women faced with the stress associated with caring for young children. The overall magnitude of the relationship between maternal depressive symptoms and short stature was somewhat less pronounced than that previously reported in South Asian studies.3032 However, comparisons are complicated by differences in study design, including growth indicators and age at measurement, control of potential confounders, and timing and method of measuring depression (e.g., major depression vs nonclinical symptoms).

In a longitudinal study of 632 urban Pakistani women, Rahman et al.32 found that a prenatal depressive disorder diagnosis predicted relative risks of short stature of 3.5 and 3.0 among infants at 6 and 12 months, respectively; corresponding relative risks of underweight at the same time points were 3.2 and 2.8. In addition, such a diagnosis was associated with a relative risk of 5 or more diarrheal disease episodes per year. Rahman et al. underscored the important role of chronic prenatal and postnatal depression in understanding infant growth and health outcomes in South Asian settings. In our own previous analyses of the present Brazilian data, regression models of height-for-age as a continuous rather than a dichotomous variable showed no evidence of an association with maternal depressive symptoms.21

In contrast to studies conducted in South Asia,3032 we found no relationship between maternal depressive symptoms and underweight; however, in those studies, analyses were not adjusted for variations in weight status attributable to stature as opposed to low weight for a given height. The higher prevalence of childhood short stature and underweight in South Asia33 than in Brazil also may reflect variability in the relative importance of proximate determinants of child growth, including regional differences in macronutrient and micronutrient intake and bioavailability and the timing and composition of complementary feeding and predominant infections; the caregiving practices of mothers with depressive symptomatology could affect the way in which these determinants are mediated or modified.

Maternal mental disorders, including symptoms of depression and anxiety, have not been shown to be related to underweight in Peru or Ethiopia.11 A Jamaican study did report a crude association of depressive symptoms with underweight, but this relationship was not significant after adjustment for maternal height and socioeconomic factors.9 As suggested by Harpham et al., there may also be cultural explanations as to why child growth delays may be more stressful for mothers in particular social contexts, including factors such as multiple child-care demands and family members’ expectations.11 High levels of demand may increase a mother’s vulnerability to depression when a child is not doing well.11 In our study, it is also possible that there was in fact a relationship between maternal depressive symptoms and child underweight, but the low prevalence of underweight in our sample did not allow enough power to detect it.

There is ample evidence of impairments in interactions between depressed parents and their children.34 In a low-income population living in the area of Santiago, Chile, anxious mother–infant attachment was related to lower weight-for-age in young childhood,35 consistent with earlier summaries of the failure to thrive literature documenting the role of feeding and non-feeding interactions in growth delays.7 Depressed caregivers may be less likely to perceive that a child is sick or respond to his or her needs, and they may be less able to coax the child to eat. Likewise, mothers with depressive symptoms may be less likely to engage in healthy feeding or sleep practices with their infant,36 less likely to breast-feed,37 and less likely to provide tactile stimulation.38 In a study with controls for dietary energy intake, infant massage led to increased weight gain among infants of depressed mothers.39

The mothers’ belief in their ability to perform well as a parent did not seem to be reflected in child undernutrition, despite our hypotheses about the potential influence of maternal self-efficacy on growth. The bulk of the literature on parenting self-efficacy has focused on its relationship to positive parenting (e.g., responsiveness, interactive behavior, and parental discipline strategies).40 Maternal self-efficacy plausibly could influence caregiving behaviors that affect proximate determinants of growth.7,41

To our knowledge, only one report has directly addressed associations between maternal self-efficacy and child growth.42 Unexpectedly, the researchers in that study found higher self-efficacy to be related to caregiving behaviors that were associated with lower weight-for-age.42 However, other caregiving behaviors, including less sensitive interactions and less positive engagement with a child,43 could be potential mediators of the association between parenting efficacy and growth and deserve further attention. The fact that we lacked data on parenting behavior limits our capacity to understand whether self-efficacy is an antecedent or consequence of parenting behaviors that lead to undernutrition. Furthermore, it is unknown whether parenting self-efficacy may be more relevant to people living in less disadvantaged circumstances than those of the mothers we studied.40

One advantage of this study was that our sanitation and SES and living conditions scales combined several aspects of these constructs. The sanitation scale included features of the neighborhood infrastructure, such as garbage collection and sewage, as well as household characteristics, such as ownership of a refrigerator and use of a water filter. The SES and living conditions scale brought together income and other aspects of wealth as reflected in possessions and type of housing. The rich detail we had on these conditions suggests that the associations observed between maternal depressive symptoms and child health were unlikely to be solely because of unmeasured indicators of socioeconomic position.

The fact that the relationship between maternal depressive symptoms and child short stature persisted after we controlled for a wide range of sociodemographic indicators gives us additional confidence in our findings. However, a limitation of this study is that our cross-sectional design focused on only a specific window of child growth between the ages of 6 and 24 months, a period during which maternal depression might not be most relevant.

Low-income urban populations in middle-income countries, similar to the sample assessed here, have been identified as being at risk for undernutrition.44 Our results from Brazil suggest that the role of maternal depressive symptoms in childhood stunting may be especially important and that programs focusing on mothers’ mental health, including prevention and treatment of depression, may result in collateral benefits for their children.


Support for this study was provided by the Maternal and Child Health Bureau (grant 5T76 MC 00001), the General Federation of Women’s Clubs of Massachusetts Catherine E. Philbin Public Health Scholarship, the Association for Women in Science Educational Foundation, the Harvard School of Public Health, and Harvard University.

We acknowledge Perspective Enterprises for donating scales and anthropometric measuring devices; UNICEF Fortaleza for providing measuring tapes; the stores Rhyme N’ Reason, Creature Feature Production, and Peacock Feather for donating toys for the children in the pilot study; and the Department of Society, Human Development, and Health at Harvard School of Public Health for data entry support.

We thank the participating families from Teresina and our interviewers, Lúcia Antônia Paiva Timbó, Jeanne Freitas Araújo, Maria do Espírito Santo Piauilino, Célia Lopes, Magda Viana da Silva, Mazé Mendes, and Marina Vieira Cardoso de Oliveira Santos, for their dedication in the field. We acknowledge Marivaldo and Arisvaldo Saraiva for extensive work on the census; Simone Ferreira for transcription of qualitative interviews; Demócrito Barreto and Rafaela Soares for technical assistance with field maps; Evôneide Gomes, Dorcas and Carlos Costa, Helotônio Carvalho, Cecelia Paz, and the staff at the Frei Damião Posto de Sáude for help with translations and coordinating pilot study activities; Eliete Batista Moura and José Ivo Pedrosa for serving as liaisons to the Universidade Federal do Piauí; Laila Caddah Ibiapina and Janaína Moura Lima for help in coordinating field work; and Alessandra de Sousa and Apolonia Nogueira for assistance in interviewer and anthropometric training.

Finally, we acknowledge the Fundação Municipal de Saúde in the municipality of Teresina, including Sílvio Mendes, Amarílis Borba, and Maria Vieira, for linkage to the Family Health Care Program; Paulo Robert Araújo Couto for creating our maps; Leandro Nery da Fonseca Coelho, Martha Fay, Yang Xu, Maria Gymour, Mack Ramsey, Lam Nguyen, and Thach Do from the Department of Health and Social Behavior; and Sid Atwood, Amy Cohen, and Keila Gomes for assistance with data management.

Human Participant Protection
This study was approved by the human subjects committee at the Harvard School of Public Health. Verbal informed consent was obtained from all participating mothers.


Peer Reviewed

P. J. Surkan designed the study, oversaw pilot work and the execution of the main study, analyzed the resulting data, and wrote the article. I. Kawachi, L. M. Ryan, L. F. Berkman, and K. E. Peterson assisted with interpretation of findings and contributed to the analysis plan and editing of the article. L. M. Carvalho Vieira assisted in supervising the fieldwork and data collection.


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