Abstract
This paper presents a new mixed demand model for measuring the value of public goods in an intertemporal optimization framework. From the specification of an indirect utility function allowing for public goods, direct demand functions for private goods are derived and estimated jointly with the Euler equation for intertemporal consumption behavior, using US data. This allows us to identify the marginal utility of private consumption and to obtain the inverse demand or shadow price of a public good, which is then related to its observed price to assess whether the public good is efficiently provided. There is evidence, though suggestive, that the public good as measured by national defense in the USA has been inefficiently provided over the past decades.
Similar content being viewed by others
Notes
Wildasin (1989) reviews the current state of research in this area and discusses possible bias in the estimation of demand for public goods resulting from distortionary taxation and other sources.
In our analysis, we are mostly concerned with public goods provided by the government; they are public goods that are privately provided.
This is a statement of the well-known Hicks’ composite commodity theorem that justifies the existence of a price index. It says that that if the prices of goods within any group always move in the same proportion, then the demand for the group as a whole has the properties identical to that for a single commodity. See Lewbel (1996) for a further discussion with evidence.
According to the Lindahl model (see Atkinson and Stiglitz 1980, for a good discussion), each person shares the burden of the marginal cost or price of the public good according to his marginal benefit received from it. Thus each person (h)’s share of the price or unit cost of the public good (\(p_{k}^{{}}\)) is \(t^{h} \times p_{k}^{{}}\) where \(t^{h}\) is the tax share of the hth person (the sum of \(t^{h}\) is 1). Then the sum of each individual’s MRS’s is equal to the sum of each person’s price share, which is equal to \(p_{k}^{{}}\). Thus if we interpret the price and the MRS in (3) as average or per capita values, respectively, our model is consistent with the representative agent model.
We consider leisure or labor supply as fixed and treat labor income as exogenous to the consumer’s choice.
Attfield (1985) endogenizes consumption by making it a function of commodity prices, using a direct utility function. For the mixed demand system, McLaren and Wong (2009) endogenize consumption expenditure for unconstrained goods by making it a function of total expenditure which is the sum of expenditures on unconstrained and constrained commodities. We employ the Euler equation with the use of GMM with relevant instruments.
The MPIGLOG is a rank 2 demand system which has the desirable property that regularity at a base level of real consumption, and at all higher levels of real consumption, can be ensured if parameter values satisfy certain simple inequality restrictions. Economic models always confront the trade-off between regularity (i.e., consistency with rational behavior) and flexibility. Researchers who insist on consistency with rational behavior are usually restricted to simple, inflexible, functional forms such as CES. Flexible functional forms, such as AIDS, QAIDS or translog, for example, can only be assured to be regular locally, in a region around a data point. In an intertemporal optimization model, it is important that regularity be satisfied over a large range of values of consumption. MPIGLOG, being effectively globally regular, provides an attractive compromise in this trade-off. In fact, we have considered several flexible rank 3 demand systems such as Banks et al. (1997) Quadratic Almost Ideal Demand System, and McLaren and Wong’s (2009) Composite Indirect Utility Function to represent the within-period conditional indirect utility function. Results of initial estimation, however, revealed that those systems violated the required concavity conditions (concave in \(C_{t}\)), which may cause serious problems when these models are used in policy analyses, particularly in situations where a model may be subjected to large shocks.
In estimation, because consumption and national defense have large units relative to commodity prices, we scale these variables; see the "Appendix A" for a detailed discussion.
A discussion of regularity conditions is provided in "Appendix B".
We considered the following set of instruments: a constant, time trend, time trend squared, disposable income lagged one period, and all commodity prices lagged one period, and national defense lagged one period.
The MPIGLOG specification (14) appears to be overparameterized for our particular dataset, which manifests itself as convergence issues in estimation, particularly in regard to the parameter \(\eta .\) While the parameter η is an additional measure of curvature, over and above the log function, it is also the case that only nonzero values of η allow for nonhomotheticity. This is somewhat puzzling, since in previous studies it is clear that \(\eta\) is identified by the demand equations alone. See, for example Cooper and McLaren (1996) where it was estimated as 0.61. When freely estimated in the full system, \(\eta\) was estimated to be greater than one, which violates regularity. Using a grid search over the interval (0,1), the value of 0.75 generated estimates of the remaining free parameters that fell within the region of regularity, ensuring that, for example, all the estimated Frisch and Marshallian own price elasticities were negative. Thus, in accordance with the trade-off discussed in footnote 7, we chose to constrain the parameter η to a value of 0.75, which leads to a slightly less flexible model, but regular, model. Even with η so restricted, the underlying model is more flexible than regular models that are typically available.
The problem arises how to test if η is not freely estimated. When η is unrestricted, we do not have convergence. Thus we tested as if η of 0.75 is unrestricted, implying that we do not have an unrestricted likelihood value relevant to the likelihood ratio test. However, since we rejected the null with the set value of η, we know we would have also rejected the null if we could estimate the model with η unrestricted, because that would have given a bigger test statistic. Hence our test, which returned a p value of zero to 5 decimal places, can be considered over-conservative.
Specifically, in our model characterized with the MPIGLOG indirect utility function (14), the parameter \(\gamma\) has the interpretation of the minimal national defense the consumer is willing to consider. With appropriate scaling of the unit of national defense, the estimated \(\gamma\) is 0.757 and z falls to almost this value in the 1970s. Thus in our formula for the shadow price in (20), the second term in the bracket, which should dominate as C grows, is divided by (\(z - \gamma\)) which is almost dividing by zero. Thus the extreme value of the estimated shadow price of national defense in the 1970s is caused by the data leading very close to a point of singularity.
The fact that the estimate of consumers’ minimal acceptable level of defense is barely below the minimum actual spending of the early 1970s is quite interesting in its own right; however, the consequence of this is a volatile estimate of the shadow price of defense in this period. Our emphasis should be on the estimated shadow price over the last 35 years.
However, while it is true there is no “market” price for defense, there is an available series on total spending on defense, and it would seem to be a reasonable assumption that this spending can be equated with the defense component of total taxation. Then using the real series on defense, and population, would appear to allow creation of a reasonably accurate proxy for price of defense.
In this study, we only considered one public good but it is possible to apply our approach to estimate shadow prices of multiple public goods. The omission of possibly relevant public goods (other than national defense) from the empirical analysis may affect or bias the estimates. We are not able to determine the direction of such bias.
References
Acemoglu D, Finkelstein A, Notowidigdo MJ (2013) Income and health spending: evidence from oil price shocks. Rev Econ Stat 95:1079–1095
Aschauer DA (1985) Fiscal policy and aggregate demand. Am Econ Rev 75:117–127
Atkinson A-S, Stiglitz J (1980) Lectures on public economics. McGraw-Hill, New York
Attfield CLF (1985) Homogeneity and endogeneity in systems of demand equations. J Econ 27:197–209
Banks J, Blundell R, Lewbel A (1997) Quadratic Engel curves and consumer demand. Rev Econ Stat 79:527–540
Barro RJ, Redlick CJ (2011) Macroeconomic effects from Government purchases and Taxes. Q J Econ 126:51–102
Bergstrom TC, Goodman RP (1973) Private demands for public goods. Am Econ Rev 63:280–296
Bergstrom TC, Rubinfeld DL, Shapiro P (1982) Micro-based estimates of demand functions for local school expenditures. Econometrica 50:1183–1205
Blundell R, Browning M, Meghir C (1994) Consumer demand and the life-cycle allocation of household expenditures. Rev Econ Stud 61:57–80
Bockstael NE, McConnell KE (1993) Public goods as characteristics of non-market commodities. Econ J 103:1244–1257
Bohm P (1972) Estimating demand for public goods: an experiment. Eur Econ Rev 3:111–130
Brookshire DS, Thayer MA, Schulze WD, D’Arge RC (1982) Valuing public goods: a comparison of survey and hedonic approaches. Am Econ Rev 72:165–177
Browning M, Deaton A, Irish M (1985) A profitable approach to labor supply and commodity demands over the life-cycle. Econometrica 53:503–543
Chavas JP (1984) The theory of mixed demand functions. Eur Econ Rev 24:321–344
Cooper RJ, McLaren KR (1992) An empirically oriented demand system with improved regularity properties. Can J Econ 25:652–668
Cooper RJ, McLaren KR (1996) A system of demand equations satisfying effectively global regularity conditions. Rev Econ Stat 78:359–364
Deaton A (1992) Understanding consumption. Oxford University Press, Oxford
Deaton AS, Muellbauer J (1980a) An almost ideal demand system. Am Econ Rev 70:312–326
Deaton A, Muellbauer J (1980b) Economics and consumer behavior. Cambridge University Press, Cambridge
Fisher D, Fleissig A, Serletis A (2001) An empirical comparison of flexible demand system functional forms. J Appl Econ 16:59–80
Gibson BB (1980) Estimating demand elasticities for public goods from survey data. Am Econ Rev 70:1069–1076
Hall RE (1978) Stochastic implications of the life cycle-permanent income hypothesis: theory and evidence. J Political Econ 86:971–987
Hall RE, Jones CI (2007) The value of life and the rise in health spending. Q J Econ 122:39–72
Hanemann WM (1991) Willingness to pay and willingness to accept: How much can they differ? Am Econ Rev 81:635–647
Hewitt D (1985) Demand for national public goods: estimates from surveys. Econ Inq 23:487–506
Kim HY (1993) Frisch demand functions and intertemporal substitution in consumption. J Money Credit Bank 25:445–454
Kim HY (2004) Commodity rates of interest and intertemporal substitution in commodity demand and consumption. Aust Econ Pap 43:228–247
Kim HY, Lee J (2001) Quasi-fixed inputs and long-run equilibrium in production: a cointegration analysis. J Appl Econ 16:41–57
Kim HY, McLaren KR, Wong KKG (2013) Empirical demand systems incorporating intertemporal consumption dynamics. Empir Econ 45:349–370
Kormendi R (1983) Government debt, government spending, and private sector behavior. Am Econ Rev 73:994–1010
Kulatilaka N (1985) Tests on the validity of static equilibrium models. J Econ 28:253–268
Lankford RH (1988) Measuring welfare changes in settings with imposed quantities. J Environ Econ Manag 15:45–63
Lewbel A (1991) The rank of demand systems: theory and nonparametric estimation. Econometrica 59:711–730
Lewbel A (1996) Aggregation without separability: a generalized composite commodity theorem. Am Econ Rev 86:524–543
McLaren KR, Wong KKG (2009) Effective global regularity and empirical modelling of direct, inverse, and mixed demand systems. Can J Econ 42:749–770
McLaughlin KJ (1995) Intertemporal substitution and λ-constant comparative statics. J Monet Econ 35:193–213
Mendelsohn RO, Olmstead SM (2009) The economic valuation of environmental amenities and disamenities: methods and applications. Annu Rev Environ Resour 34:325–347
Moschini G, Moro D (1994) Autocorrelation specification in singular equation systems. Econ Lett 46:303–309
Muellbauer J (1975) Aggregation, income distribution and consumer demand. Rev Econ Stud 42:525–543
Perkins GM (1977) The demand for local public goods: elasticities of demand for own price, cross prices, and income. Natl Tax J 30:411–422
Ramey VA (2011) Identifying government spending shocks: it’s all in the timing. Q J Econ 126:1–50
Ramey V, Shapiro M (1997) Costly capital reallocation of government spending. Carnegie-Rochester Conf Ser Public Policy 48:145–194
Varian HR (2010) Microeconomic analysis, 3rd edn. Norton, New York
Wildasin DE (1989) Demand estimation for public goods. Distortionary taxation and other sources of bias. Reg Sci Urban Econ 19:353–379
Wong KKG, Park H (2007) The use of conditional cost functions to generate estimable mixed demand systems. Am J Agric Econ 89:273–286
Author information
Authors and Affiliations
Corresponding author
Additional information
Publisher’s Note
Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.
The authors would like to thank the referees for helpful comments and suggestions to improve the paper.
Appendices
Appendix A: Scaling of z and/or C for the MPIGLOG indirect utility function in Sect. 3.1
Estimation of the budget share system and Euler equation involves different units of measurement for the variables under consideration. In our analysis, the public good in particular has large units relative to private goods, which causes convergence problems. Specifically, in line with the representative agent framework, private consumption is measured in per capita value by dividing it by population. We use national defense as a public good in empirical analysis. While it is common to express national defense in per capita term (see, e.g., Barro and Redlick 2011; Ramey 2011), this procedure is not valid because of its nature of a public good, especially joint consumption. If not expressed in per capita term, national defense has a large unit. A common procedure is to scale down those variables that have large units of measurement. It can be shown that, with appropriate scaling of some parameters, parameter estimates are invariant to this scaling.
To do this, we introduce scaling factors for the public good and consumption, k1 and k2, and rewrite (14) as
where \(\tilde{z}_{t}^{{}} = k_{1} z_{t}^{{}} ,\hat{C}_{t}^{{}} = k_{2} C_{t} ,\bar{\kappa } = (k_{1} )_{{}}^{\omega } k_{2} \kappa ,\) and \(\tilde{\gamma } = k_{1} \gamma .\) Re-expressing (A1),
This is invariant to scaling because of the constant factors right out of the intertemporal criterion function. Thus scaling of either or both of C and z does not affect the optimal solution, provided the parameters \(\kappa\) and \(\gamma\) are scaled accordingly. In our estimation, we only allow z to be scaled.
Appendix B: Regularity conditions for the MPIGLOG indirect utility function in Sect. 3.1
Provided that the conditions on parameters noted in Sect. 3.1 are satisfied, the MPIGLOG indirect utility function (14) is regular with respect to the within-period properties over the region \(C_{t}^{{}} z_{t}^{\omega } > \kappa P_{1} \left( {{\mathbf{p}}_{t} } \right)\) and \(z_{t}^{{}} > \gamma .\) Intertemporal properties of MPIGLOG with respect to the concavity of \(C_{t}^{{}}\) can be examined from (18). If \(\eta \ge 1,\) the MPIGLOG indirect utility function (14) is convex in \(C_{t}^{{}} .\) If \(0 < \eta \le 0.5,\) on the other hand, it is concave in \(C_{t}^{{}}\). If \(0.5 < \eta < 1,\) the MPIGLOG indirect utility function is likely to be concave in \(C_{t}^{{}}\) for certain combinations of data and parameters. Intertemporal optimization, however, requires the concavity of the intertemporal utility function (4). To this end, we can use the relevant expressions in (14), (15), and (16) to evaluate the concavity condition in (6). For \(\zeta > 0\) and \(0 < \eta < 0.5,\) with \(C_{t}^{{}} z_{t}^{\omega } > \kappa P_{1} \left( {{\mathbf{p}}_{t} } \right)\) and \(z_{t}^{{}} > \gamma\), the concavity of the indirect utility function (14) will preserve the concavity of the intertemporal utility function. However, if \(\zeta > 0\) and \(0 < \eta < 1,\) an additional sufficient condition is \(\zeta > 1 + (0.5/\eta ).\) If \(\eta\) is in the range \(0.5 < \eta < 1,\) then the Box–Cox transformation has to be sufficiently concave to ensure overall concavity.
Rights and permissions
About this article
Cite this article
Kim, H.Y., McLaren, K.R. & Wong, K.K.G. Valuation of public goods: an intertemporal mixed demand approach. Empir Econ 59, 2223–2253 (2020). https://doi.org/10.1007/s00181-019-01734-0
Received:
Accepted:
Published:
Issue Date:
DOI: https://doi.org/10.1007/s00181-019-01734-0