Abstract
This paper addresses the impact of productive government expenditure on income inequality using a dataset of 80 countries over the period of 1980–2015. It incorporates the conflicting predictions implied by alternative growth models on this issue. While the neoclassical model suggests that productive government expenditure will reduce long-run income inequality, the corresponding endogenous growth model suggests the opposite. We examine this proposition, by considering both the aggregate Gini coefficient, and the income shares of quintiles. The results obtained using the Gini coefficients provide compelling support for the contrasting impacts of government investment on income inequality, suggested by the underlying theoretical models. These findings are supported, albeit somewhat more weakly by the regressions employing the quintile data. Our general conclusion is that government investment has a mixed effect on income inequality, a conclusion consistent with previous studies.
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Notes
This has generated a voluminous literature focusing on both the theoretical aspects and the empirical evidence. An extensive survey of the theoretical literature is provided by Agénor, (2013). Most of this literature focuses on government investment in isolation. Chen (2006) shows that the growth effects of government spending are enhanced if simultaneously the government chooses its optimal combination of consumption and investment. The overwhelming consensus of the empirical evidence is that infrastructure contributes positively and significantly to output, although its exact magnitude is subject to debate. See Bom and Ligthart (2014) for an exhaustive review of the empirical literature.
Lopez (2004) is one of the few papers to analyze systematically the effect of infrastructure on both growth and income inequality. Using panel data he finds that infrastructure raises the growth rate, while reducing income inequality.
This contrast is evident by comparing the distributional effects of an increase in TFP for the basic one-sector endogenous growth model in García-Peñalosa and Turnovsky (2006) with those obtained assuming the neoclassical production function by Turnovsky and García-Peñalosa (2008). In a related study, Bouza and Turnovsky (2012) employ a two-sector dependent economy model to consider the impact of a foreign transfer on wealth and income inequality. Among their findings they show that if the transfer is allocated to the nontraded sector, it will tend to reduce inequality. Since most government investment tends to be nontraded, this result exhibits a parallel to the implications of the one-sector Ramsey model being considered here.
There is also an extensive literature focusing on the impact of trade globalization on income inequality. This relationship generally tends to be more tenuous, a results that we also tend to find.
This term was introduced by Caselli and Ventura (2000). It is a direct extension of the representative agent model in which all agents have equal unimpeded access to all markets.
While productive government spending directly enhances private productivity, it differs from a pure productivity shock in that it entails a claim on the economy’s resources.
As in Barro’s (1990) seminal work, productive government expenditure is introduced as a flow. Insofar as it represents infrastructure, it may be more appropriately introduced as a public capital good, as in Futagami et al. (1993). Since the role of the underlying model is primarily to facilitate the interpretation of the empirical results, the simpler “flow” formulation suffices for this purpose. Also, because we do not introduce the underlying public capital, we prefer to characterize the government expenditure as “productive” rather than as “investment”, although both terms may be used.
In that case there would also be a problem of compatibility since G(t) would be growing indefinitely, while measures of inequality, such as the Gini coefficient, are bounded.
With the return to capital being specified by \(r(t)\), this measure of relative income is net of depreciation.
It is important to point out that wealth inequality evolves only during the transition of \(L(t)\). In the event that \(L(t)\) jumps instantaneously to steady state, \(\omega (t) \equiv 1\) and wealth inequality remains unchanged, \(\sigma_{k} (t) = \sigma_{k,0}\); see (11).
While this explanation, which reflects the underlying pattern of the transitional dynamics, is helpful in understanding the empirical results, we should keep in mind that it is derived under restrictive conditions, most significantly that of a Cobb–Douglas technology. There are several natural modifications to this stylized representation that weaken the dichotomy, although it still remains. These are discussed in an expanded version of this paper.
March 2017 version.
IMF provides an index (base = 2010) for the Terms of Trade.
In specifying (21) we should note the following issue. While the underlying theory expresses inequality in terms of the coefficient of variation, among the alternative summary measures of inequality proposed by Atkinson (1970), the Gini coefficient is the closest and hence the most convenient for empirically implementing the theory.
It is well known that including a lagged dependent variable in dynamic panel models introduces certain issues with respect to their estimation. These are alluded to in the discussion of the empirical estimates in Sect. 7, below.
We do not report the dummy variables for alternative sources, country, or time, referred to in Sect. 5.
Robust standard errors are clustered by country in the fixed effects estimation.
See Wooldridge (2005).
Arellano and Bond (1991).
The number of instruments is also shown: the number of instruments is relatively small with respect to the number of observations. Too many instruments can weaken the Hansen test of the instruments´ joint validity [see Roodman (2009a) for instance]. We have employed xtabond2 command in the estimation, as developed by Roodman (2009b).
This regression is implemented using the command “rreg” in the statistical package Stata.
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Acknowlgements
We thank Branko Milanovic for helpful discussions concerning the data. Stephen J. Turnovsky's research was supported in part by the Van Voorhis endowment at the University of Washington. Iñaki Erauskin acknowledges financial support from Programa de Movilidad del Personal Investigador del Departamento de Educación, Política Lingüística y Cultura del Gobierno Vasco and Programa de apoyo a los grupos del sistema universitario vasco del Departamento de Educación, Política Lingüística y Cultura del Gobierno Vasco (Grupo de investigación IT885-16). This paper was written in part while Iñaki Erauskin was visiting the University of Washington. The remaining errors and omissions are entirely the responsibility of the authors.
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Appendices
Appendix 1
Appendix 1 provides some of the details supporting the results reported in the text.
1.1 Properties of aggregate transitional matrix
The local transitional dynamics of the aggregate economy summarized by Eq. 7a, b is of the form:
where \( \begin{gathered} a_{{11}} = (1 - g)\left[ {F_{K} + F_{G} G_{K} } \right] - \frac{{(1 - g)}}{\eta }\left[ {F_{{LK}} + F_{{LG}} G_{K} } \right](1 - L) - \delta ;\,\, \hfill \\ \,a_{{12}} = (1 - g)\left[ {F_{L} + F_{G} G_{L} } \right] - \frac{{(1 - g)}}{\eta }\left[ {(F_{{LL}} + F_{{LG}} G_{L} )(1 - L) - F_{L} } \right] \hfill \\ b_{1} = - (1 - g)[F_{{KK}} + F_{{KG}} G_{K} ] > 0;\,\,\,\,b_{2} = - (1 - g)[F_{{KL}} + F_{{KG}} G_{L} ] < 0 \hfill \\ F(K,L,G) = AK^{\alpha } L^{{1 - \alpha }} G^{\nu } ;\,\,\,\,G = \left( {gAK^{\alpha } L^{{1 - \alpha }} } \right)^{{\frac{1}{{1 - \nu }}}} \hfill \\ \Omega = L\left\{ {(1 - \gamma )\left[ {\frac{1}{{1 - L}} - \frac{{(1 - \alpha )}}{{1 - \nu }}} \right] - \frac{{\gamma \eta L}}{{1 - L}}} \right\}^{{ - 1}} ;~\,~\kappa = \left( {1 - \gamma } \right)\frac{\alpha }{K} > 0;\,\,\,\, \hfill \\ \end{gathered} \).
Using the fact that the underlying production function is Cobb–Douglas, together with the steady-state relationships (9a)–(9c), these expressions may be simplified as follows:
One can readily show that the determinant of the transitional matrix: \(\Omega (a_{11} b_{2} - a_{21} b_{1} ) < 0\), implying that the aggregate transitional dynamic path, summarized by the relationship between \(L(t)\) and \(K(t)\) is a saddlepoint, the stable path of which is described by
With \(a_{12} > 0\), the slope of the relationship (A.2) depends upon \({\text{sgn}} (\mu - a_{11} )\). To determine this we first solve for the stable eigenvalue, \(\mu < 0\), in terms of its trace, T, and its determinant, \(\Delta\), to obtain \(\mu = \left( {{1 \mathord{\left/ {\vphantom {1 2}} \right. \kern-\nulldelimiterspace} 2}} \right)\left( {T - \sqrt {T^{2} - 4\Delta } } \right)\). Using this expression it is straightforward to show that \(\mu - a_{11} < 0\) if and only if \((\kappa a_{11} + b_{1} ) > 0\). Using the above to evaluate this expression, we see that
where \(\varepsilon \equiv (1 - \gamma )^{ - 1}\) is the intertemporal elasticity of substitution. Clearly (A.3) is an extremely mild restriction and will hold for any plausible set of parameters.
1.2 Derivation of Eqs. (10) and (11)
Following the procedure outlined in Sect. 3.2.1 yields the following equation in \(\dot{k}_{i} (t)\):
To obtain the bounded solution we linearize (A.4) around the steady state. In doing so we note that (9b) and the homogeneity of the production function, imply
Using the equilibrium optimality conditions (9c) and labor market clearance then yields
from which we immediately infer the restrictions
Using (A.6), the bounded solution to (A.4) is
where \(\omega (t) \equiv 1 + \left( {\frac{{(1 - g)F_{L} (\tilde{K},\tilde{L},G)}}{{\tilde{K}}}\left[ {\frac{{L(t) - \tilde{L}}}{{1 - \tilde{L}}}} \right]} \right)\left( {\frac{{(1 - g)F_{L} (\tilde{K},\tilde{L},G)}}{{\tilde{K}}}\frac{1}{\eta }\left[ {1 - (1 + \eta )\tilde{L}} \right] - \mu } \right)^{ - 1}\).
Substituting (A.5) this expression simplifies to
Using (2a) and (2b) in conjunction with the steady-state conditions we obtain:
And recalling (9c) we derive:
Appendix 2
2.1 List of countries
-
(1)
Developed OECD countries: Australia, Austria, Belgium, Canada, Czech Republic, Denmark, Finland, France, Germany, Greece, Ireland, Italy, Republic of Korea, Netherlands, Norway, Portugal, Spain, Sweden, United Kingdom, United States.
-
(2)
Other high income countries: Croatia, Cyprus, Estonia, Israel, Lithuania, Malta, Poland, Slovak Republic, Slovenia, Switzerland.
-
(3)
Middle income countries Argentina, Belarus, Belize, Botswana, Brazil, China, Colombia, Costa Rica, Dominican Republic, Ecuador, Iran, Kazakhstan, Macedonia FYR, Malaysia, Mexico, Namibia, Paraguay, Peru, Romania, Russian Federation, South Africa, Thailand, Turkey, Uruguay, Venezuela.
-
(4)
Middle-Low income countries: Bangladesh, Cape Verde, Djibouti, Egypt, El Salvador, Ethiopia, Guatemala, Honduras, India, Indonesia, Lesotho, Mongolia, Morocco, Nigeria, Pakistan, Sri Lanka, Tajikistan, Ukraine.
-
(5)
Low income countries: Central African Republic, Comoros, Congo, Nepal, Senegal, Tanzania, Uganda.
2.2 List of countries WIID
-
(1)
High income developed OECD countries: Austria, Belgium, Cyprus, Czech Republic, Denmark, Finland, France, Germany, Greece, Ireland, Italy, Malta, Netherlands, New Zealand, Norway, Portugal, Spain, Sweden, Switzerland, United Kingdom.
-
(2)
Other high income countries: Croatia, Estonia, Lithuania, Poland, Slovak Republic, Slovenia.
-
(3)
Middle income countries: Argentina, Belize, Brazil, Colombia, Costa Rica, Dominican Republic, Ecuador, Macedonia FYR, Mexico, Paraguay, Peru, Turkey, Uruguay, Venezuela.
-
(4)
Middle-Low income countries: El Salvador, Guatemala, Honduras.
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Turnovsky, S.J., Erauskin, I. Productive government expenditure and its impact on income inequality: evidence from international panel data. Rev World Econ 158, 331–364 (2022). https://doi.org/10.1007/s10290-021-00433-2
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DOI: https://doi.org/10.1007/s10290-021-00433-2