Abstract
We estimate efficiency and TFP growth for two measures of congestion and two measures of the monetary value of congestion for the largest 88 contiguous cities in the U.S. over the period 1982–2007. Using stochastic frontier analysis we find that the efficiency scores for congestion and the associated ranking of cities is sensitive to the measure of congestion. In contrast, the efficiency scores and rankings are robust for the two measures of the monetary value of congestion. Most importantly, for the most valid measure of congestion and both measures of the monetary value of congestion, we find that average TFP growth over the study period is characterized by an upward trend. This is an encouraging sign even though in all three cases growth is only zero or slightly less than zero at the end of the study period. We therefore conclude that policies which have been used towards the end of the study period such as providing incentives to carpool and encouraging employers to offer flexi-time and telecommuting arrangements appear to have been effective and should be implemented more widely.
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Notes
We use urban areas and cities interchangeably. Urban areas are determined by U.S. Census demographic criteria and typically tend to be associated with a distinct city.
See Fernández et al. (2005) for a detailed explanation on how to manipulate the output-oriented efficiency for the transformed output to obtain environmental efficiency.
We justify the choice of inputs and outputs in the data section using, among other things, the results of Hausman-Wu tests, which test whether endogeneity has a significant effect on the consistency of the estimated parameters. We also provide in the data section details of the measure of congestion which the TTI initially used together with details of the measure of congestion which they now favor. Why we favor one measure of the monetary value of congestion over the other is explained in the data section.
We thank an anonymous referee for recommending that we calculate public transit elasticities outside the sample mean for quintiles of the population size distribution.
The measures of the monetary value of congestion are at 1982 prices. The TTI values of person travel time and commercial vehicle time are not based on the prevailing wage rate but on the perceived valuation of delay. We therefore deflate the TTI measure of the monetary value of congestion using the CPI. Our second measure is based on the wage rate so we deflate it using the GDP deflator.
Our second measure of the monetary value of congestion is therefore only based on the value of person travel time and overlooks the value of commercial vehicle time and the value of wasted fuel.
Because of data availability issues the climatic data for 2007 is assumed to apply throughout the study period.
As of March 2007, there were 88 HOV facilities in California, followed by 83 in Minnesota, 41 in Washington State, 35 in Texas and 21 in Virginia.
To plot the densities in Figs. 3 and 5 we use the Gaussian density and obtain the bandwidth h using the Sheather and Jones (1991) solve-the-equation plug-in-approach. When estimating the kernel densities to avoid bias problems near the boundary, the reflection method, as described by Silverman (1986) and Scott (1992), is used.
The correlation and Spearman rank correlation coefficients for the efficiency scores from models 1 and 2 are both around 0.46. For the efficiency scores from models 2–4, all the correlation and Spearman rank correlation coefficients are above 0.91. The Wilcoxon rank-sum (Mann-Whitney) test rejects the null of equality of the efficiency distributions for models 1 and 2 at the 1 % level with a z-statistic of 7.68. This suggests that the two efficiency distributions are drawn from different populations. As was expected, the null for the Wilcoxon rank-sum test is not rejected for pairs of efficiency distributions from models 2–4.
To construct Figs. 4 and 6 we use bivariate Gaussian kernels and the bandwidths are calculated using the solve-the-equation plug-in approach for a bivariate Gaussian kernel à la Wand and Jones (1994). In Fig. 4, the efficiency scores are normalized relative to the mean. As was expected, the contour plots of the efficiency scores from model 1 against the efficiency scores from model 2, 3 or 4 are indistinguishable so only the contour plot of the efficiency scores from model 1 against the scores from model 4 is presented.
The basis for the income categorization is simply the ranking of cities according to mean real personal income per head over the study period.
The correlation and Spearman rank correlation coefficients for the rates of TFP growth from models 1 and 2 are 0.01 and −0.04, respectively. The correlation and Spearman rank correlation coefficients for the rates of TFP growth from model 1 and models 3–4 range from 0.00 to 0.01 and −0.05 − (−0.07), respectively. The correlation and Spearman rank correlation coefficients for the TFP growth rates from models 2–4 are much higher and range from 0.40 to 0.47 and from 0.86 to 0.94, respectively.
The Wilcoxon rank-sum tests of the null of equality of TFP growth distributions from models 1 to 4 are rejected if the test involves the distribution from model 1. The other tests accept the null.
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Acknowledgments
The authors would like to thank two anonymous referees and Tom Weyman-Jones for constructive comments on an earlier draft of this paper. We also acknowledge the comments from participants at the 2011 European Workshop on Efficiency and Productivity Analysis (EWEPA) and the 2011 Kuhmo Nectar Conference on Transportation Economics.
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Appendix
Appendix
See Table 3.
1.1 Test of the concavity condition
Microeconomic theory assumes that the form of an input distance function, D I , satisfies a particular curvature property i.e. D I is linearly homogeneous and concave in inputs x. The symmetry property and homogeneity of degree one in x of D I implies that only K(K − 1)/2 elements of the Hessian matrix, H(x), where \(H(x)\equiv \frac{\partial ^{2}D_{I}}{\partial x\partial x^{\prime }}\), are linearly independent. Applying the arguments of Diewert and Wales (1987), the Hessian of the input distance function with respect to x can be calculated as follows:
where: \(\widehat{ex}\) is a diagonal matrix with estimated input elasticities, ex k for \(k=1,\ldots ,K-1\), on the leading diagonal and zeros elsewhere; ex is a vector of estimated input elasticities, \(ex_{k}; \mathbf{B}\) is the matrix of second order coefficients on the input terms in the translog function.
Concavity of the input distance function in x requires that the Hessian matrix is negative semidefinite. Whether the Hessian matrix is negative semidefinite can be verified from the sign pattern of the principal minors of the Hessian. The necessary and sufficient conditions for D I to be concave are as follows: all the odd-numbered principal minors of the Hessian must be non-positive and all the even-numbered principal minors must be non-negative. At the sample mean with mean corrected data, the Hessian is given by
where \(\widehat{\beta }\) is a diagonal matrix with estimated input elasticities, β k for \(k=1,\ldots ,K-1\), on the leading diagonal and zeros elsewhere, and β is a vector of estimated input elasticities, β k . The Stata code for the concavity test is available from the corresponding author on request.
1.2 Efficiency scores
See Table 4.
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Glass, A.J., Kenjegalieva, K. & Sickles, R.C. How efficiently do U.S. cities manage roadway congestion?. J Prod Anal 40, 407–428 (2013). https://doi.org/10.1007/s11123-012-0288-9
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DOI: https://doi.org/10.1007/s11123-012-0288-9
Keywords
- Congestion
- Monetary value of congestion
- Panel data
- Stochastic frontier analysis
- Input distance function
- TFP