Abstract
In this paper, comparable surveys from 1994 to 2008 are used to show that two geographically and culturally connected nations, Canada and the United States (US), have starkly contrasting violence rates among children and youth and that this cross-country violence gap has emerged among those as young as 2 years old for hitting, 4 years old for bullying, and 12 years old for fighting. Such early life differences remain important as children grow up. The US-Canada violence gaps do not appear to reduce as personal and family characteristics are adjusted for, for example, race, family structure, poverty, and region. Policies in areas most relevant for childhood development, including maternity or parental leave, health care, and child care, are compared across the two countries to identify potential areas where more attention may be paid to improve children’s outcomes.
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Notes
Studies on violence among younger children within countries are also scarce. See Vlachou et al. (2011) for a comprehensive review.
Besides prevalence, it might also be interesting to examine the intensity of violent behavior, that is, the number of times a child has hit, bullied, or fought. Fu et al. (2013) and ? show that prevalence and intensity of bullying victimization may exhibit different trends. However, Canada-US comparative measures of intensity of violence cannot be constructed in this research, because the response categories related to intensities, as shown in Table 1, are not comparable across the two countries.
The number of observations is greater than the number of children because these are pooled cross-sections and some children appear repeatedly in the survey cycles. All analysis was done for the samples with and without repeated observations. The results with repeated observations are presented in this paper. Due to space limit, the results without repeated observations are not presented here but are available upon request.
Starting from Cycles 5 of the NLSCY, cross-sectional weights for the original cohort are not available anymore. For those observations, longitudinal weights are used instead. The NLSCY also provides bootstrap weights to reflect the complex survey design. However, to facilitate direct comparison with the US results, the Canadian results are not bootstrapped because bootstrap weights are not supplied in the CNLSY79.
The NLSY79 over-samples disadvantaged US families, which leads to the concern that this over-sampling will result in upward biases in the US violence rates. The cross-sectional weights supplied in the CNLSY79 and employed in all analysis throughout the paper take into account this over-sampling. To be cautious, however, all analysis is also carried out without those over-sampled observations and the results are robust.
Though not the focus of this paper, the gender gap in disruptive behavior has been well-documented and causes for this gap have been studied in the literature. Bertrand and Pan (2013) review that literature and provide evidence that home environment matters in addition to biological differences between genders.
Suppose, during the initial period, period 0, child A’s probability of being violent is 20 %, thus her probability of being non-violent is 80 %. \(T = \left (\begin {array}{ll} 0.3 & 0.7\\ 0.1&0.9 \end {array} \right )\) is the transition matrix for child A. This means that, if child A was violent in period t−1, there is a 30 % chance that she will again be violent and a 70 % chance that she will not be violent in period t. Similarly, if child A was not violent in period t−1, her probabilities of being violent and non-violent in period t are 10 % and 90 %, respectively. Thus, child A’s probabilities of being violent and non-violent in period 1 are 0.14 and 0.86, respectively, since \((0.2, 0.8)\times \left ( \begin {array}{ll} 0.3 & 0.7\\ 0.1&0.9 \end {array} \right ) = (0.14, 0.86)\). Child A’s probabilities of being violent and non-violent in period 2 are 0.128 and 0.872, respectively, since \((0.14, 0.86)\times \left (\begin {array}{ll} 0.3 & 0.7\\ 0.1&0.9 \end {array} \right ) = (0.128, 0.872)\). If this calculation is continued until the 12th period, then child A’s probabilities of being violent and non-violent stabilize at 0.125 and 0.875, respectively, and will not change any further into the future. Child A’s ergodic probability of violence is thus 0.125. Suppose child B’s initial probabilities of being violent and non-violent are 40 % and 60 %, different from child A’s, while her transition probabilities are the same as child A’s. By the same token, child B’s violence and non-violence probabilities in period 1 can be calculated as \((0.4, 0.6)\times \left ( \begin {array}{ll} 0.3 & 0.7 \\ 0.1&0.9 \end {array} \right ) = (0.18, 0.82)\), which are different from child A’s probabilities in period 1. However, by period 12, child B’s violence and non-violence probabilities also converge to (0.125, 0.875) and will no longer change. Thus, this example shows that child A and child B’s ergodic probabilities of violence are the same due to their identical transition probabilities despite the fact that they start with different initial probabilities of violence.
Some Canadian provinces with relatively smaller population bases are grouped together.
Three Canadian cities are classified as central-cities here, Montreal in Quebec, Toronto in Ontario, and Vancouver in BC. US central city boundaries are defined by the US Census Bureau. For details, please refer to Appendix 6 of the NLSY 79 Codebook.
Poverty lines are \(\frac {1}{2} \times \) median equivalent family income in each country - the Luxembourg Income Study definition. Family equivalent income is defined as family income divided by the square root of family size to capture within-family economies of scale, the Luxembourg Income Study equivalence scale. For US, the poverty line is calculated using the Current Population Survey 2003. For Canada, it is calculated using the 2003 Survey of Labor Income and Dynamics.
Family equivalent income is defined as family income divided by the square root of family size to capture within-family economies of scale, the Luxembourg Income Study equivalence scale. Income is expressed in Purchasing-Power-Parity and inflation adjusted 2003 real US dollars.
Following Burton and Phipps (2007), the idea is to capture economies of scale within family with respect to time, similar to the concept of equivalent family income. The calculation of “equivalent adult time available” is (# of parents × 112 − total parental weekly work hours)/(square root of family size), where 112 is total number of hours per week, 168 (24 hours/day × 7 days), minus sleep hours per week 56 (8 hours/day × 7 days).
A sizeable portion, around 15 %, of US observations had missing values for their family income, as it had to be added up across a number of items to be comparable to the Canadian income variable, which is a mother-reported before tax and after transfer income measure. Instead of dropping those US observations with missing income values, an additional independent variable is included indicating such cases.
As mentioned in Section 2, regressions that do not include repeated observations yield similar results and are available upon request.
The IE, however, could not be estimated for fighting due to insufficient number of observations in some age-period cells.
It should be noted that the Age-Period-Cohort results presented here provide the average US-Canada differences in children’s violence rates holding constant age, period, and cohort. However, the age, period, and cohort effects may be different in each country, which is not allowed in the above estimates. Ideally, it will be useful to estimate the Age-Period-Cohort models either separately for Canada and US or by interacting the age, period, and cohort variables with the US dummy. This can then help address the question of whether different age, period, and cohort effects between Canada and US may account for some of the observed cross-country differences in children’s violence rates. However, these additional analyses are not carried out due to sample size consideration. As described in Section 2, the analytical sample used in this research is Canadian and US children born to those women who were between 14 and 21 years old as of December 31, 1978. The fertility cycle of this cohort of women determines that the number of children in some age-period cells is relatively small. More specifically, there are fewer older children towards the beginning and fewer younger children towards the end of the period under study, 1994–2008. Thin age-period cells prevent reliable estimates from being obtained. Currently when the Age-Period-Cohort models are estimated with the Canada-US pooled data, the Intrinsic Estimator, could not be estimated for fighting due to insufficient number of observations in some age-period cells - see footnote 16 and the notes to Appendix Table 15 and Figs. 6, 9, and 12. Further splitting the cells by country would exacerbate this problem.
Specifically, Canadian 10–17 year olds are asked the following question: “During the past 12 months, about how many times were you questioned by the police about anything that they thought you did?” US youth who are at least 15 years old are asked the following question: “Since date of last interview, have you ever been convicted of any charges other than a minor traffic violation?” Probit regressions are estimated with these indicators of encounter with the criminal justice systems as dependent variables. The right-hand-side variables include indicators of early-life violent behaviors and the same set of explanatory variables as those in Tables 10 to 12. These regression results are not reported due to space limit, but are available upon request.
Currie and Stabile (2007) find negative connections between childhood behavioral problems and future test scores and schooling attainment for both Canada and US.
In Canada, provincial governments are in charge of legislations on job-protected maternity or parental leave and the federal government funds the income compensation. With some variations, the duration of job-protected leave in most provinces has been in keeping with the federal UI/EI rules. Thus, the discussion here will focus on the federal legislation.
See Phipps (2006) for a thorough discussion of the evolution of Canadian maternity and parental benefits.
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Acknowledgments
I would like to thank Yulia Kotlyarova, Lars Osberg, Shelley Phipps, Barbara Wolfe, Kuan Xu, and Jim Marshall for their help. I thank the Canadian Labour Market and Skills Researcher Network, the Social Sciences and Humanities Research Council of Canada, and the Office of Research Services of the University of Regina for funding. This work was carried out at Statistics Canada’s Atlantic Research Data Center and Saskatchewan Research Data Center.
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Appendix
Appendix
Age Effects for Hitting, Gaussian Link, without Region Dummies, 2-6 Year Olds. Data Source: NLSCY, NLSY79, and CNLSY79, 1994–2008. The results are similar if including region dummies or using the Probit link function. Coefficients obtained from the Constrained Generalized Linear Model are centered to be comparable to those obtained from the Intrinsic Estimator
Age Effects for Bullying, Gaussian Link, without Region Dummies, 4–11 Year Olds. Data Source: NLSCY, NLSY79, and CNLSY79, 1994–2008. The results are similar if including region dummies or using the Probit link function. Coefficients obtained from the Constrained Generalized Linear Model are centered to be comparable to those obtained from the Intrinsic Estimator
Age Effects for Fighting, Gaussian Link, without Region Dummies, 12–17 Year Olds. Data Source: NLSCY, NLSY79, and CNLSY79, 1994–2008. The Intrinsic Estimator could not be estimated for fighting due to insufficient number of observations in some age-period cells. The results are similar if including region dummies or using the Probit link function. Coefficients obtained from the Constrained Generalized Linear Model are centered to be comparable to those obtained from the Intrinsic Estimator
Period Effects for Hitting, Gaussian Link, without Region Dummies, 2–6 Year Olds. Data Source: NLSCY, NLSY79, and CNLSY79, 1994–2008. The results are similar if including region dummies or using the Probit link function. Coefficients obtained from the Constrained Generalized Linear Model are centered to be comparable to those obtained from the Intrinsic Estimator
Period Effects for Bullying, Gaussian Link, without Region Dummies, 4–11 Year Olds. Data Source: NLSCY, NLSY79, and CNLSY79, 1994–2008. The results are similar if including region dummies or using the Probit link function. Coefficients obtained from the Constrained Generalized Linear Model are centered to be comparable to those obtained from the Intrinsic Estimator
Period Effects for Fighting, Gaussian Link, without Region Dummies, 12–17 Year Olds. Data Source: NLSCY, NLSY79, and CNLSY79, 1994–2008. The Intrinsic Estimator could not be estimated for fighting due to insufficient number of observations in some age-period cells. The results are similar if including region dummies or using the Probit link function. Coefficients obtained from the Constrained Generalized Linear Model are centered to be comparable to those obtained from the Intrinsic Estimator
Cohort Effects for Hitting, Gaussian Link, without Region Dummies, 2–6 Year Olds. Data Source: NLSCY, NLSY79, and CNLSY79, 1994–2008. The results are similar if including region dummies or using the Probit link function. Coefficients obtained from the Constrained Generalized Linear Model are centered to be comparable to those obtained from the Intrinsic Estimator
Cohort Effects for Bullying, Gaussian Link, without Region Dummies, 4–11 Year Olds. Data Source: NLSCY, NLSY79, and CNLSY79, 1994–2008. The results are similar if including region dummies or using the Probit link function. Coefficients obtained from the Constrained Generalized Linear Model are centered to be comparable to those obtained from the Intrinsic Estimator
Cohort Effects for Fighting, Gaussian Link, without Region Dummies, 12–17 Year Olds. Data Source: NLSCY, NLSY79, and CNLSY79, 1994–2008. The Intrinsic Estimator could not be estimated for fighting due to insufficient number of observations in some age-period cells. The results are similar if including region dummies or using the Probit link function. Coefficients obtained from the Constrained Generalized Linear Model are centered to be comparable to those obtained from the Intrinsic Estimator
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Zhang, L. Violence in Early Life: A Canada-US Comparison. Child Ind Res 8, 299–346 (2015). https://doi.org/10.1007/s12187-014-9234-x
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DOI: https://doi.org/10.1007/s12187-014-9234-x