The unprecedented large-scale rural-to-urban migration in China has left many rural children living apart from their parents. In this study, we examine the impact of parental migration on the nutritional status of young children in rural areas. We use the interaction terms between wage growth, by gender, in provincial capital cities and initial village migrant networks as instrumental variables to account for migration selection. Our results show that parental migration has no significant effect on the height of children, but it improves their weight. We provide suggestive evidence that the improvement in weight may be achieved through increased access to tap water in migrant households.
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A focus on children’s nutrition is important because poor nutrition outcomes early in life have long-term consequences for children (Grantham-McGregor et al. 2007; Glewwe and Miguel 2008). For example, multiple studies demonstrate that children’s health and nutritional status have a sizable and statistically significant positive impact on later educational outcomes in many developing countries, including Ghana (Glewwe and Jacoby 1995), Guatemala (Maluccio et al. 2009), Kenya (Miguel and Kremer 2004), and Pakistan (Alderman et al. 2011). Benefits are not limited to educational outcomes: Longitudinal studies in developing countries also show evidence of improved cognitive measures later in life (Grantham-McGregor et al. 2007). Furthermore, stunting before the age of 5 can lead to less formal employment or lower wages later in life (Walker et al. 2011; Maluccio et al. 2009; Carba et al. 2009).
They also find that children who accompany migrants are healthier using broader measures of health status. Further, in a separate study, Stillman et al. (2012) find migration increases obesity among 3- to 5-year-olds, and the impacts are likely to be driven by dietary change.
BMI stands for body mass index, calculated by dividing a person’s body weight in kilograms by her or his height in meters squared [weight (kg)/height (m)2]. BMI z-scores are used like weight-for-height z-scores but also control explicitly for age.
Gibson et al. (2011) use a lottery to credibly identify Tonga–New Zealand migration.
Chen (2006) uses earlier rounds of the CHNS to show that child health among older children is not affected by father absence in the past month. She also finds that caloric intakes increase when fathers are away. We do not study caloric intakes in this paper partly because adequate caloric intake is not likely to affect child height, which requires both a hygienic environment and adequate. We are also concerned that the variable of the change in caloric intake over 2–4 years suffers from systematic measurement error.
China’s labor market experienced dramatic changes during the 1990s as the number of rural residents moving to urban areas for employment grew rapidly. Estimates using the 1 % sample from the 1990 and 2000 rounds of the Population Census and the 1995 1 % population survey suggest that the intercounty migrant population grew from just over 20 million in 1990 to 45 million in 1995 and to 79 million by 2000 (Liang and Ma 2004). These figures likely underreport the true scale of migration because they do not account for seasonal or shorter term migration (Cai et al. 2008).
There may be differences in outcomes related to the gender of the parent who leaves. If the mother specifically leaves, as she is the primary care giver the father may not act as a perfect substitute. We do not differentiate by gender of the migrant parent in part because migration by mother alone is relatively rare and because we lack statistical power to differentiate migration by gender.
See http://www.cpc.unc.edu/projects/china for details.
The nine provinces are Henan, Hubei, Heilongjiang, Liaoning, Shandong, Guizhou, Jiangsu, and Guangxi, Hunan.
See http://www.who.int/childgrowth/en/ and http://www.who.int/growthref/en/ for details on the growth reference data, which are consistent with one another.
Unfortunately, the CHNS lacks information on remittances back to households from migrants. Therefore, we measure the net effect of migration, meaning the variable is the net effect of all effects of migration on the household (including any remittances received).
Children with an HAZ below −2 are considered stunted.
We include the self-reported values of the following types of assets in the calculation of total asset values: productive assets (transportation means, farm machinery, draft animals, commercial equipment), household electrical appliances (radio, tape recorder, VCR, black/white television, color television, washing machine, refrigerator, air conditioner, sewing machine, electric fan, computer, camera, microwave oven, electric rice cooker, pressure cooker, telephone, VCD, or DVD). Only 4.6 % of households reported no assets. When we take the logarithm of the asset value, we assign a value of 1 to those households.
Although we would prefer to build up the network from an earlier round, such as the 1993 round, the reasons for absence were recorded only from the 1997 round onward, so we cannot use earlier waves to construct the instrumental variables. For villages in Liaoning province, which were added to the survey in 2000, the 2000 migration rates are used to measure the migrant network.
In the sample, each village on average has about 48 women and 49 men in this age group.
The influence of controlling for household wealth and demographic changes on the relationship between migration and nutritional status is ambiguous. On the one hand, parental emigration is likely to correlate positively with the number senior household members because small children are more likely to live with a grandparent in migrant than nonmigrant households (de Brauw and Mu 2011). As the number of senior female household members appears positively correlated with children’s nutritional outcomes (Table 2), the estimated migration effect would be larger without household demographic variables. Households may increase out-migration in response to negative wealth shocks (Giles and Yoo 2007), constituting a negative correlation between migration and household assets. Whereas household wealth is positively associated with child nutrition, without asset variable the estimated effect of emigration would be smaller.
Results are not reported, but available from the authors upon request.
Limited by the sample size, we are not able to precisely estimate heterogeneous impacts along the gender or poverty dimension.
The results estimated from the restricted sample show that the coefficient on the change in parent migration status for the change in weight-for-age z-scores is 0.201, significant at 10 % level. The coefficients are 0.034 and −0.033 for the regressions of the change in HAZ scores and BMIZ, respectively, and they are not statistically significant.
Another potential measure of household investment in hygiene is whether there are excreta around the house, as observed and reported by the survey numerators. We also used this variable as a dependent variable and do not find a statistically significant coefficient on migration.
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We are grateful to Nicole Garcia and Feng Huang for great research assistance. For very useful comments, we thank three anonymous referees, Joyce Chen, Kalena Cortes, Ranjita Misra, Hiroshi Ono, and seminar participants at the 2012 Allied Social Science Associations Annual Conference in Chicago, the Asian Studies Forum at Texas A&M University, and the 2013 Population Association of America Conference in New Orleans.
Responsible editor: Klaus F. Zimmermann
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Mu, R., de Brauw, A. Migration and young child nutrition: evidence from rural China. J Popul Econ 28, 631–657 (2015). https://doi.org/10.1007/s00148-015-0550-3
- Rural China