Abstract
This paper estimates production functions for the 48 contiguous U.S. states over the 1970–2000 period employing recently developed panel estimators that tackle simultaneously parameter heterogeneity, cross-sectional dependence and non-stationarity. The findings suggest that labor, private capital and, when controlling for cross-sectional dependence, average schooling years exert a positive and significant effect on state income. In contrast, the income effect of public capital stock is negative. The state-specific coefficients indicate that this effect likely stems from the negative elasticity of income with respect to public capital in the states located primarily in the Snow Belt region.
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Notes
It should be also noted that there is an upsurge of interest in estimating region-wide production functions outside the USA. See, for example, Merriman (1990), Yamano and Ohkawara (2000) and Shioji (2001) for Japan; Looney and Frederiksen (1981) for Mexico; Nijkamp (1986) for the Netherlands; Cutanda and Paricio (1994), Mas et al. (1996), Cantos et al. (2005) and Bajo-Rubio and Díaz-Roldán (2005) for Spain.
Conditional convergence refers to convergence toward the steady-state income level after accounting for initial conditions.
A number of papers investigate the relationship between human capital and productivity across geographic areas outside the USA. See, inter alia, Rivera and Currais (2004), Gumbau-Albert and Maudos (2006) and Ramos et al. (2010) for Spanish regions; Di Giacinto and Nuzzo (2006), Di Liberto (2008), Bronzini and Piselli (2009) and Costantini and Destefanis (2009) for Italian regions; Gundlach (1997), Zhang and Zhang (2003) and Fleisher et al. (2010) for Chinese provinces; and Audretsch and Keilbach (2004) for German regions.
In this case, the empirical specification is obtained by ignoring public and human capital and performing a linear transformation of the variables in Eq. (3a) to implement the required returns to scale restriction. Thus, labor productivity becomes the dependent variable, and the private capital/labor ratio is used as a regressor. The empirical specification (in log form) becomes:
$$\begin{aligned} y_{ it }-l_{ it }= & {} a_{i}(k_{ it }-l_{ it })+u_{ it }\\ u_{ it }= & {} \varphi _{i}+\lambda _{i}^{\prime }f_{t}+\varepsilon _{ it }. \end{aligned}$$Labor income shares are calculated following the procedure proposed by Gollin (2002). First, the wage and salary income of employees is imputed as labor income. Then, the average labor income of employees is calculated, and the same average labor income is imputed to the self-employed. The sum of the two is used as a measure of total labor income. Dividing total labor income by total income provides an estimate of the labor income share (lis). The capital income share (cis) is then determined residually as 1-lis. Data on total income, employees’ wages and employment, and the non-farm proprietors’ employment are available from the BEA Web site (http://www.bea.gov/regional/).
Engelbrecht (2002) and Papageorgiou (2003), who examine a world sample, distinguish between primary and post-primary education. However, we believe that the distinction between secondary and tertiary education is more appropriate for the U.S. states, where primary education attainment is almost universal, and there is some cross-state variation in secondary education, but most variation concerns tertiary education completion (descriptive statistics for all variables are available upon request). Also, the above distinction is more relevant, since we investigate a highly skill-intensive economy like the USA.
Bond and Eberhardt (2013) explain that the differenced year dummies are extracted from the pooled regression in first differences because non-stationary variables and unobservable common factors can severely bias the estimates in the pooled level regressions.
For instance, \(\hat{a}_{\mathrm{AMG}}=1/N\sum \nolimits _{i=1}^{N} a_{i}\).
We obtain qualitatively similar results when we include a trend in the unit root test specifications (the results are available upon request).
In Table 7 in “Appendix”, we estimate Eq. (3a) using standard spatial econometric techniques (SAR and SAC models) in an attempt to compare our baseline results to those obtained by Dall’erba and Llamosas-Rosas (2015). Similarly to them, we find that private and public capital have a positive direct effect on the income of the state where they are allocated (although the effect of public capital is imprecisely estimated in SAR). In contrast to them, we find that average years of schooling also exert a significant positive direct effect on income. More important, however, for the purposes of the present study are the estimates of the indirect effects, which reflect spatial spillovers. The indirect effects are highly insignificant in both models, thereby suggesting that the cross-sectional dependence evidenced in Table 1 goes beyond what geographic proximity can explain. This, in turn, justifies the use of the CCEMG and AMG estimators, which allow for a more general common factor specification of cross-sectional dependence compared to spatial econometric techniques.
We also estimated the regressions without state-specific time trends but the parameter estimates remained largely unaltered (the results are available upon request).
The proportion of the population with at least a high school degree is the most general metric of human capital stock provided by Frank (2009) as it also includes those holding a college degree. The pairwise correlation between average schooling years and the secondary school completion rate is 0.74. It should be noted here that population is typically characterized by lower education attainment than the labor force. For instance, according to the 2000 U.S. Census, the nationwide proportion of population with at least high school education was 76.9 %, while the corresponding share of the labor force was 84.6 %.
The negative effect of secondary school attainment also accords with the empirical literature on growth and human capital. Savvides and Stengos (2009, p. 108) write: “[s]tudies that treat human capital as a direct input to the production function have shown that human capital accumulation exerts an insignificant or sometimes even negative effect on growth...”.
Note that the row elements of a spatial weights matrix measure the impact on a particular state by all other states, while the column elements of a spatial weights matrix display the impact of a particular state on all other states. Consequently, row normalization has the effect that the impact on each state by all other states is equalized. So, the standardized version of spatial weights is \(W_{ ij }\left( \omega _{ ij }{/}\sum _{i=1}^{j}\omega _{ ij }\right) \).
We have tested for fixed versus random effects via Hausman tests and have always concluded in favor of fixed effects.
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Acknowledgments
We thank especially the Associate Editor and two anonymous referees for their very constructive remarks and suggestions. The paper has also benefited greatly from comments received by S. Karkalakos. We also thank K. Christ for the provision of the public capital data used in this paper. The usual disclaimer applies.
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Appendix
Appendix
Column (1) of Table 7 reports the estimation results using specification (3a) modified in the form of a spatial autoregressive (SAR) model (LeSage and Pace 2009, chapter 1). In effect, we estimate:
where both i and \(j=1,{\ldots },48\) denote a state, and \(t=1,\ldots ,31\) a time period. The SAR model posits that the dependent variable is a function of the dependent variable observed in neighboring states and a set of observed local characteristics. Specifically, rho is the spatial autoregressive coefficient and \(W_{ ij }\) is the lag weight matrix of dimension \(48\times 48\) (there are 48 states in the sample), which measures the geographical distance between any pair of states i, j. We assume that the elements of the weight matrix are inversely related to the distance between states, i.e., \(\omega _{ ij }=(1/d_{ ij })\), where \(d_{ ij }\) is the Euclidean distance between the centroids of states i and j. Also, we posit that the maximum distance where spillovers take place is equal to the \(1{\mathrm{st}}\) quartile of the distribution of all distance pairs between any pair of U.S. states, so that spillovers do not occur among very distant states. This structure of weights is such that states contribute to the geographical spillovers proportionally to their distance, so that weights penalize more the distant states. Spatial weights are row-standardized, so that for each state the sum of weights equals one.Footnote 15 Thus, column (1) posits that real personal income in each state depends on that of its neighboring states up to a threshold distance, at a rate which decreases proportionally to the distance of each neighboring state from the state in question. We believe that this is a more flexible assumption than that of Dall’erba and Llamosas-Rosas (2015), who only allow for spillovers from the four closest neighbors. We include time and state fixed effects to allow for time trend and state-specific unobserved characteristicsFootnote 16 and compute bootstrapped cluster-robust standard errors according to Pace and LeSage (2009).
Column (2) reports the estimates of the spatial autocorrelation (SAC) model (LeSage and Pace 2009, chapter 2), which incorporates both spatial effects of the dependent variable and spatial effects among the error terms. In effect, we estimate:
Thus, SAC nests the SAR model and assumes that real personal income in each state depends on (1) real personal income of its neighbors as defined above and (2) a set of unobserved, common characteristics omitted from the model. The estimated values of rho and lambda are the spatial parameters of the spatially lagged dependent variable and spatially autocorrelated error terms, respectively. We estimate both SAR and SAC models with maximum likelihood. According to the results, the rho parameter is statistically insignificant in both specifications. In contrast, the lambda parameter is highly significant in the SAC model (column 2). In that sense, the SAC model seems to be a better choice than SAR in our setup and provides evidence for the existence of significant spatial effects in the U.S. states working primarily through the error terms.
The upper panel of Table 7 reports the coefficient estimates of the total effects of the explanatory variables (labor, private capital, total public capital stock and average schooling years), while the lower panels report the corresponding direct and indirect estimates. The total effects measure the impact of a unit change in the above variables in all states on personal income in state i, averaged over all states. The direct effects measure the impact of a unit change in state i on personal income in state i, again averaged over all states. The indirect impact estimates are the difference of total and direct effects, which economists usually refer to as spillovers and depend on geographical distance in our specifications. When considering the direct effects, we observe that in our setup all four state inputs exert a sizeable, positive and statistically significant impact on the respective state incomes. These findings are in line with Dall’erba and Llamosas-Rosas (2015) who also find positive direct effects for private capital, public capital and human capital (although the latter is statistically insignificant). The presence or absence of significant spillovers depends on whether the indirect effects that arise result in statistically significant estimates. We observe that all indirect effects are statistically insignificant in both specifications, thereby casting doubt on the geographical proximity as the main driver for the presence of spillovers across U.S. states. Thus, observable production factors causing spillovers across U.S. states do not seem to be related to distance between states.
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Benos, N., Mylonidis, N. & Zotou, S. Estimating production functions for the US states: the role of public and human capital. Empir Econ 52, 691–721 (2017). https://doi.org/10.1007/s00181-016-1092-6
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DOI: https://doi.org/10.1007/s00181-016-1092-6