Abstract
We extend a recently developed generalized local polynomial estimator into a semiparametric smooth coefficient framework to estimate a generalized cost function. The advantage of the generalized local polynomial approach is that we can simultaneously choose the degree of polynomial for each continuous nonparametric regressor and the bandwidths via data-driven methods. We provide estimates of scope economies from the joint production of microloans and microdeposits for a dataset of Microfinance Institutions from over 50 countries. Our approach allows analysis on all Microfinance Institutions rather than only those offering just microloans. Moreover, the smooth coefficient estimator provides a general interface in which to account for both direct and indirect environmental factors. We find substantial scope economies in general, of about 10 % at the median, as well as evidence that economies of scope vary across the type of services and country in which the MFIs operate, suggesting key insights into policy prescriptions.
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Notes
Throughout the paper, we use microdeposits (microsavings) to mean voluntary microdeposits since mandatory savings that MFIs require are a part of some of the lending technology associated with solidarity groups and village banks.
See Du et al. (2013) for an approach to impose linear homogeneity in this setting.
Note that this generalized formulation nests the familiar local constant least squares problem when \(p_s = 0, \,\forall s\)
$$\begin{aligned} \frac{1}{nh} \sum _{i=1}^n \left[ Y_i - \sum _{\ell =0}^L a_\ell X_{\ell i} \right] ^2 {\mathcal {K}}\left( \frac{Z_i-z}{h} \right) \end{aligned}$$or the local linear least-squares problem when \(p_s = 1, \,\forall s\)
$$\begin{aligned} \frac{1}{nh} \sum _{i=1}^n \left[ Y_i - \sum _{\ell =0}^L a_\ell X_{\ell i} - \sum _{\ell =0}^L \sum _{s=1}^{S} b_{\ell s}(Z_{si}-z_s)X_{\ell i} \right] ^2 {\mathcal {K}}\left( \frac{Z_i-z}{h} \right) . \end{aligned}$$We also explore the robustness of our primary results by selecting bandwidths that are optimal for estimation of economies of scale. We find that, in general, our empirical results are qualitatively consistent when we smooth over scale.
For a detailed comparison of this with other available datasets see Mersland (2009).
Distribution of MFIs by country is presented in the “Appendix”. Comparison with other publicly available data shows that these data have more observations from Latin America.
Differences in funding costs may also stem from differences in inflation across countries and different risk premiums. Inflation is, however, to a large extent taken care of in the dataset since all amounts are converted into US dollars.
Since the cost function is homothetic in input prices, we can always normalize by one of the input prices. Thus, while we have three inputs, only two of them enter into our analysis.
Multistarts are the number of different trials used to calculated the minimum of the least-squares cross-validation function. Given the nonlinearity of this function with respect to multiple bandwidths, it is good practice to use numerous multistarts to avoid obtaining bandwidths indicative of a local minimum as opposed to the global minimum.
We note that with the use of flexible estimation methods that theoretical consistency may be sacrificed. In our case, this entails the estimated cost function satisfying given axioms of producer theory, notably monotonicity of the cost function in both outputs, loans and deposits. The percentage of observations where our estimated cost function is non-monotonic for loans is 6.7 % when we include controls and roughly 3 % with no controls. However, for deposits, across both models roughly 50 % of the estimated cost function derivatives are negative. What drives this behavior is a single, smooth coefficient term which dominates the expression—the interaction between loans and deposits. The coefficient on this term is largely negative, and when multiplied by the amount of loans in the derivative of cost w.r.t. deposits, we have a largely negative term. This matters because many other terms in this derivative are zero, since deposits is mostly zero for almost 75 % of the observations. While this is certainly an important issue to explore, we leave it for future research to combine the approach detailed here with constrained nonparametric methods.
We also considered augmented variants of the 45\(^{\circ }\) plots presented in Fig. 1 that differentiate between institutions in our sample that offer only loans and those that jointly offer savings and loans, to assess whether there are large distributional differences between scope estimates across these groups. We omit these plots since we were not able to identify distinct differences across the scope estimate distributions across these two groups. Note that this merely indicates that the heterogeneity we identify in our estimates of scope economies affects scope economies across these two groups similarly.
As a comparison, we consider our standard scope measure for only this subset of 178 countries and find that, while there is a wider interquartile range of our estimates, the qualitative conclusion from our main results is unchanged.
It is also worth noting that most of these studies measure output by the number of active borrowers (or clients), while we use the volume of loans and deposits.
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We thank Daniel Henderson, participants at the 2013 Western Economic Association International conference, and seminar participants at Hebei Finance University for helpful feedback.
Appendix
Appendix
Distribution of MFIs by country is presented in Table 10. Comparison with other publicly available data shows that these data have more observations from Latin America, perhaps because they needed external funds.
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Delgado, M.S., Parmeter, C.F., Hartarska, V. et al. Should all microfinance institutions mobilize microsavings? Evidence from economies of scope. Empir Econ 48, 193–225 (2015). https://doi.org/10.1007/s00181-014-0861-3
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DOI: https://doi.org/10.1007/s00181-014-0861-3