Abstract
We analyze behavior of food prices in India during the last decade at a disaggregate level. Systematic decomposition shows that eggs, meat, fish, milk, cereals, and vegetables are the main contributors to food inflation. Fruits and vegetables showed a much higher short-term volatility in prices. All the major contributors possess a higher weight in the consumption basket, indicating that the weight of a commodity has much larger bearing on its overall contribution to food inflation, as compared to other factors such as base effect or percentage change in prices (inflation). The inflation-volatility patterns reveal that the commodities that have higher income elasticity of demand but have limited processing and storage facilities, such as fruits and vegetables, are characterized by higher volatility. Econometric analysis shows that while cereal and edible oil prices appear to be mainly driven by supply-side factors such as production, wage rates, and minimum support prices, for pulses, the effects of supply and demand factors appear almost equal. On the other hand, prices of eggs, meat, fish, milk, and fruits and vegetables appear to be driven mainly by demand-side factors. Price projections show that the eggs-meat-fish-milk group shows the highest increase because of the higher income elasticities of demand and the rapid increase in India’s per capita income in the recent years.
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Notes
There has been a slowdown in food inflation since August 2014, mainly due to the higher base effect and the slowdown in food and fuel prices globally. However, understanding long-term trends is still important for informed policy making. In addition, food inflation has started inching up again since April 2015.
MPCE is a proxy for income spent on the particular group of commodities. The variable is lagged in order to avoid reverse causality.
In an ongoing study by the authors, it has been found that the major contributing commodities are broadly similar across major states in India, with minor differences.
For a few of the commodities, although the models are the best fitted as per diagnostic statistics, the coefficient restrictions are not satisfied.
Inflation here refers to magnitude of commodity’s inflation and not the commodity’s contribution to overall inflation, which depends also upon the weight of the commodity and the base effect. This distinction is important while interpreting the results in this section. Also, we have used volatility measure computed using the ratio method in order to get one single measure of volatility for each commodity.
Because current expenditure is a function of current price, which is the focus of our inquiry.
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The authors would like to extend special thanks to the anonymous reviewer of the journal for very insightful suggestions/comments. Special thanks to the International Food Policy Research Institute (IFPRI) South Asia office and the Institute of Economic Growth (IEG) for the support provided. Thanks are also due to the participants of various conferences at IFPRI, IEG, Indian Statistical Institute and Centre for Economic and Social Studies, where this paper was presented, for their comments/suggestions. The usual disclaimer applies.
Appendices
Appendix 1: ARCH/GARCH models
The generalized autoregressive conditional heteroscedasticity GARCH(p,q) model is given by
where \(\alpha_{i} \ge 0, \, \beta_{j} \ge 0,\alpha_{i} + \beta_{j} < 1\),\(\varepsilon_{t}^{{}}\) is the error term, \(h_{t}\) is the conditional variance in t, and \(\alpha_{i} {\text{ and }}\beta_{j}\) are the parameters (to be estimated). If p = q = 1, then this model reduces to a GARCH(1,1) model, \(h_{t} = \alpha_{0} + \alpha_{i} \varepsilon_{t - 1}^{2} + \beta_{j} h_{t - 1}^{{}}\). Variance \(h_{t}\) in period t is conditional on (1) the size of the residual in t − 1 and (2) \(h_{t - 1}\), the conditional variance in the previous period. Thus, \(h_{t}\) is large if \(\varepsilon_{t - 1}^{2}\) is large, and vice versa.
1.1 TARCH model (threshold ARCH model)
TARCH is an extension of the GARCH model; it allows the effects of positive and negative shocks (to prices in this case) to have different effects on volatility. The standard TARCH model for a one-period lag is given by
where \(d_{t - 1} = 1\) if \(\varepsilon_{t - 1}^{{}} < 0\) and \(d_{t - 1} = 0\) if \(\varepsilon_{t - 1}^{{}} \ge 0\). In addition, (1) \(h_{t} = \alpha_{0} + (\alpha_{1} + \lambda_{1} )\varepsilon_{t - 1}^{2} + \beta_{1} h_{t - 1}^{{}}\) in case of a negative error, and (2) \(h_{t} = \alpha_{0} + \alpha_{1} \varepsilon_{t - 1}^{2} + \beta_{1} h_{t - 1}^{{}}\) otherwise; and \(\alpha_{1} \ge 0, \, \beta_{1} \ge 0\), but there is no restriction on \(\lambda_{1}\).
1.2 EGARCH model (exponential GARCH model)
In the GARCH and TARCH models, there are binding coefficient restrictions such as\(\alpha_{i} \ge 0, \, \beta_{j} \ge 0{\text{ and }}\alpha_{i} + \beta_{j} < 1\). A less restrictive model is the exponential GARCH (EGARCH) model, given by
In this model, both \(\alpha_{i}\) and \(\beta_{j}\) can have negative values since the dependent variable is \(\ln (h_{t} )\) and not \(h_{t}\). The effect of \(\varepsilon_{t - 1}^{{}}\) on \(\ln (h_{t} )\) is equal to \(\alpha_{1} + \lambda_{1}\) when \(\varepsilon_{t - 1}^{{}} \ge 0\) and equal to \(- \alpha_{1} + \lambda_{1}\) when \(\varepsilon_{t - 1}^{{}} < 0\).
Appendix 2: Volatility of commodity subgroups, 2005–2014, all India
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Sekhar, C.S.C., Roy, D. & Bhatt, Y. Food inflation and volatility in India: trends and determinants. Ind. Econ. Rev. 53, 65–91 (2018). https://doi.org/10.1007/s41775-018-0017-z
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DOI: https://doi.org/10.1007/s41775-018-0017-z