\chapter{GMM estimation}
\label{chap:gmm}

\section{Introduction and terminology}
\label{sec:gmm-intro}

The Generalized Method of Moments (GMM) is a very powerful and general
estimation method, which encompasses practically all the parametric
estimation techniques used in econometrics. It was introduced in
\cite{hansen82} and \cite{hansen-singleton82}; an excellent and
thorough treatment is given in chapter 17 of
\cite{davidson-mackinnon93}.

The basic principle on which GMM is built is rather straightforward.
Suppose we wish to estimate a scalar parameter $\theta$ based on
a sample $x_1, x_2, \ldots, x_T$.  Let $\theta_0$ indicate the ``true''
value of $\theta$. Theoretical considerations (either of statistical
or economic nature) may suggest that a relationship like the following
holds:
\begin{equation}
  \label{eq:simple}
  E \left[ x_t - g(\theta) \right] = 0 \Leftrightarrow \theta =
  \theta_0 ,
\end{equation}
with $g(\cdot)$ a continuous and invertible function. That is to say,
there exists a function of the data and the parameter, with the
property that it has expectation zero if and only if it is evaluated
at the true parameter value.  For example, economic models with
rational expectations lead to expressions like (\ref{eq:simple}) quite
naturally.

If the sampling model for the $x_t$s is such that some version of the
Law of Large Numbers holds, then
\[
  \bar{X} = \frac{1}{T} \sum_{t=1}^T x_t \convp g(\theta_0) ;
\]
hence, since $g(\cdot)$ is invertible, the statistic
\[
  \hat{\theta} = g^{-1}(\bar{X}) \convp \theta_0 ,
\]
so $\hat{\theta}$ is a consistent estimator of $\theta$. A different
way to obtain the same outcome is to choose, as an estimator of
$\theta$, the value that minimizes the objective function
\begin{equation}
  \label{eq:obj-simple}
  F(\theta) = \left[ \frac{1}{T} \sum_{t=1}^T (x_t  - g(\theta)) \right]^2 =
  \left[ \bar{X} - g(\theta) \right]^2 ;
\end{equation}
the minimum is trivially reached at $\hat{\theta} = g^{-1}(\bar{X})$,
since the expression in square brackets equals 0.

The above reasoning can be generalized as follows: suppose $\theta$ is
an $n$-vector and we have $m$ relations like
\begin{equation}
  \label{eq:GMMgeneral}
  E \left[ f_i(x_t, \theta) \right] = 0 \quad\textrm{for\ } i=1 \ldots
  m ,
\end{equation}
where $E[\cdot]$ is a conditional expectation on a set of $p$
variables $z_t$, called the \emph{instruments}. In the above simple
example, $m=1$ and $f(x_t, \theta) = x_t - g(\theta)$, and the only
instrument used is $z_t = 1$. Then, it must also be true that
\begin{equation}
  \label{eq:oc}
  E \left[ f_i(x_t, \theta) \cdot z_{j,t} \right] = E \left[ f_{i,j,t}(\theta) \right] = 
  0 \quad\textrm{for\ } i=1 \ldots
  m \quad\textrm{and \ } j=1 \ldots p;
\end{equation}
equation (\ref{eq:oc}) is known as an \emph{orthogonality condition},
or \emph{moment condition}. The GMM estimator is defined as the
minimum of the quadratic form
\begin{equation}
  \label{eq:obj-general}
  F(\theta, W) = \bar{\mathbf{f}}' W \bar{\mathbf{f}},
\end{equation}
where $\bar{\mathbf{f}}$ is a $(1 \times m\cdot p)$ vector holding the
average of the orthogonality conditions and $W$ is some symmetric,
positive definite matrix, known as the \emph{weights} matrix. A
necessary condition for the minimum to exist is the order condition $n
\le m \cdot p$. 

The statistic
\begin{equation}
  \label{eq:gmmestimator}
  \hat{\theta} = \argmin_{\theta} F(\theta, W)
\end{equation}
is a consistent estimator of $\theta$ whatever the choice of $W$.
However, to achieve maximum asymptotic efficiency $W$ must be
proportional to the inverse of the long-run covariance matrix of the
orthogonality conditions; if $W$ is not known, a consistent estimator
will suffice.

These considerations lead to the following empirical strategy:
\begin{enumerate}
\item Choose a positive definite $W$ and compute the
  \emph{one-step} GMM estimator $\hat{\theta}_1$. Customary choices
  for $W$ are $I_{m\cdot p}$ or $I_{m} \otimes (Z'Z)^{-1}$.
\item Use $\hat{\theta}_1$ to estimate $V(f_{i,j,t}(\theta))$ and use its
  inverse as the weights matrix. The resulting estimator
  $\hat{\theta}_2$ is called the \emph{two-step} estimator.
\item Re-estimate $V(f_{i,j,t}(\theta))$ by means of $\hat{\theta}_2$ and
  obtain $\hat{\theta}_3$; iterate until convergence. Asymptotically,
  these extra steps are unnecessary, since the two-step estimator is
  consistent and efficient; however, the iterated estimator often has
  better small-sample properties and should be independent of the
  choice of $W$ made at step 1. 
\end{enumerate}

In the special case when the number of parameters $n$ is equal to the
total number of orthogonality conditions $m \cdot p$, the GMM
estimator $\hat{\theta}$ is the same for any choice of the weights
matrix $W$, so the first step is sufficient; in this case, the
objective function is 0 at the minimum. 

If, on the contrary, $n < m \cdot p$, the second step (or successive
iterations) is needed to achieve efficiency, and the estimator so
obtained can be very different, in finite samples, from the one-step
estimator. Moreover, the value of the objective function at the
minimum, suitably scaled by the number of observations, yields
\emph{Hansen's J statistic}; this statistic can be interpreted as a
test statistic that has a $\chi^2$ distribution with $m \cdot p -n $
degrees of freedom under the null hypothesis of correct specification.
See Davidson and MacKinnon (\citeyear{davidson-mackinnon93}, section
17.6) for details.

In the following sections we will show how these ideas are
implemented in gretl through some examples.

\section{GMM as Method of Moments}
\label{sec:gmm-as-mom}

\emph{This section draws from a kind contribution by Alecos
  Papadopoulos, whom we thank.}

A very simple illustration of GMM can be given by dropping the ``G'',
via an example of the time-honored statistical technique known as
``method of moments''; let's see how to estimate the parameters of a
gamma distribution, which we also used as an example for ML estimation
in section~\ref{sec:ml-gamma}.

Assume that we have an i.i.d. sample of size $T$ from a gamma
distribution. The gamma density can be parameterized in terms of the
two parameters $p$ (shape) and $\theta$ (scale), both real and
positive.\footnote{In section \ref{sec:ml-gamma} we used a slightly
  different, perhaps more common, parametrization, employing
  $\theta = 1/\alpha$. We are switching to the shape/scale
  parametrization here for the sake of convenience.}  In order to
estimate them by the method of moments, we need two moment conditions
so that we have two equations and two unknowns (in the GMM jargon,
this amounts to exact identification). The two relations we need are
\[
  E(x_i) = p \cdot \theta \qquad V(x_i) = p \cdot \theta^2
\]

These will become our moment conditions; substituting the finite
sample analogues of the theoretical moments we have
\begin{eqnarray}
  \label{eq:mm-ex-mean}
  \bar{X} & = & \hat{p} \cdot \hat{\theta} \\
  \label{eq:mm-ex-var}
  V(x_i) & = & \hat{p} \cdot \hat{\theta}^2
\end{eqnarray}

Of course, the two equations above are easy to solve analytically,
giving $\hat{\theta} = \frac{\hat{V}}{\bar{X}}$ and $\hat{p} =
\frac{\bar{X}}{\hat{\theta}}$, ($\hat{V}$ being the sample variance of
$x_i$), but it's instructive to see how the \cmd{gmm} command will
solve this system of equations numerically.

We feed gretl the necessary ingredients for GMM estimation
in a command block, starting with \texttt{gmm} and ending with
\texttt{end gmm}. Three elements are compulsory within a \texttt{gmm}
block:
\begin{enumerate}
\item one or more \texttt{orthog} statements
\item one \texttt{weights} statement
\item one \texttt{params} statement
\end{enumerate}
The three elements should be given in the stated order.

The \texttt{orthog} statements are used to specify the orthogonality
conditions.  They must follow the syntax
\begin{code}
  orthog x ; Z
\end{code}
where \texttt{x} may be a series, matrix or list of series and
\texttt{Z} may also be a series, matrix or list. Note the structure of
the orthogonality condition: it is assumed that the term to the left
of the semicolon represents a quantity that depends on the estimated
parameters (and so must be updated in the process of iterative
estimation), while the term on the right is a constant function of the
data.

The \texttt{weights} statement is used to specify the initial
weighting matrix and its syntax is straightforward. 

The \texttt{params} statement specifies the parameters with respect to
which the GMM criterion should be minimized; it follows the same logic
and rules as in the \texttt{mle} and \texttt{nls} commands.

The minimum is found through numerical minimization via BFGS (see
chapters~\ref{chap:numerical} and~\ref{chap:mle}).  The
progress of the optimization procedure can be observed by appending
the \option{verbose} switch to the \texttt{end gmm} line.

\bigskip

Equations \ref{eq:mm-ex-mean} and \ref{eq:mm-ex-var} are not yet in
the ``moment condition'' form required by the \cmd{gmm} command. We
need to transform them and arrive at something looking like $E(e_{j,i}
z_{j,i}) = 0$, with $j=1 \ldots 2$. Therefore, we need two
corresponding observable variables $e_1$ and $e_2$ with corresponding
instruments $z_1$ and $z_2$ and tell gretl that $\hat{E}(e_j
z_j) = 0$ must be satisfied (where we used the $\hat{E}(\cdot)$
notation to indicate sample moments).

If we define the instrument as a series of ones, and set $e_{1,i} =
x_i - p\theta$, then we can re-write the first moment condition as
\[
\hat{E}[(x_i - p\theta) \cdot 1] = 0.
\]
This is in the form required by the \cmd{gmm} command: in the
required input statement ``\verb|orthog e ; z|'', \texttt{e} will be
the variable on the left (defined as a series) and \texttt{z} will the
variable to the right of the semicolon. Since $z_{1,i} = 1$ for all
$i$, we can use the built-in series \texttt{const} for that.

For the second moment condition we have, analogously,
\[
\hat{E}\left\{\left[(x_i - \bar{X})^2 - p\theta^2\right] \cdot 1\right\} = 0,
\]
so that by setting $e_{2,i} = (x_i - \bar{X})^2 - p\theta^2$ and $z_2
= z_1$ we can re-write the second moment condition as $\hat{E}[e_{2,i}
\cdot 1] = 0$.

The weighting matrix, which is required by the \cmd{gmm} command, can
be set to any $2 \times 2$ positive definite matrix, since under exact
identification the choice does not matter and its dimension is
determined by the number of orthogonality conditions. Therefore, we'll
use the identity matrix.

Example code follows:
\begin{scode}
# create an empty data set 
nulldata 200

# fix a random seed
set seed 2207092

#generate a gamma random variable with, say, shape p = 3 and scale theta = 2
series x = randgen(G, 3, 2)  

#declare and set some initial value for parameters p and theta
scalar p = 1				
scalar theta =1
								
#create the weight matrix as the identity matrix
matrix W = I(2)

#declare the series to be used in the orthogonality conditions
series e1 = 0				
series e2 = 0				

gmm
    scalar m = mean(x)
    series e1 = x - p*theta
    series e2 = (x - m)^2 - p*theta^2
    orthog e1 ; const
    orthog e2 ; const
    weights W
    params p theta
end gmm
\end{scode}

The corresponding output is
\begin{code}
Model 1: 1-step GMM, using observations 1-200

             estimate   std. error     z     p-value 
  ---------------------------------------------------
  p          3.09165     0.346565    8.921   4.63e-19 ***
  theta      1.89983     0.224418    8.466   2.55e-17 ***

  GMM criterion: Q = 4.97341e-28 (TQ = 9.94682e-26)
\end{code}


If we want to use the unbiased estimator for the sample variance, we'd
have to adjust the second moment condition by substituting
\begin{code}
series e2 = (x - m)^2 - p*theta^2
\end{code}
with
\begin{code}
scalar adj = $nobs / ($nobs - 1)
series e2 = adj * (x - m)^2 - p*theta^2
\end{code}
with the corresponding slight change in the output:
\begin{code}
Model 1: 1-step GMM, using observations 1-200

             estimate   std. error     z     p-value 
  ---------------------------------------------------
  p          3.07619     0.344832    8.921   4.63e-19 ***
  theta      1.90937     0.225546    8.466   2.55e-17 ***

  GMM criterion: Q = 2.80713e-28 (TQ = 5.61426e-26)
\end{code}
One can observe tiny improvements in the point estimates, since both
moved a tad closer to the true values. This, however, is just a
small-sample effect and not something you should expect in larger
samples.

\section{OLS as GMM}
\label{sec:gmm-ols}

Let us now move to an example that is closer to econometrics proper:
the linear model $y_t = x_t \beta + u_t$.  Although most of us are
used to read it as the sum of a hazily defined ``systematic part''
plus an equally hazy ``disturbance'', a more rigorous interpretation
of this familiar expression comes from the \emph{hypothesis} that the
conditional mean $E(y_t|x_t)$ is linear and the \emph{definition} of
$u_t$ as $y_t - E(y_t|x_t)$.

From the definition of $u_t$, it follows that $E(u_t|x_t) = 0$. 
The following orthogonality condition is therefore available:
\begin{equation}
  \label{eq:oc-ols}
  E \left[ f(\beta) \right] = 0 ,
\end{equation}
where $f(\beta) = (y_t - x_t \beta) x_t$. The definitions given in
section \ref{sec:gmm-intro} therefore specialize here to:
\begin{itemize}
\item $\theta$ is $\beta$;
\item the instrument is $x_t$;
\item $f_{i,j,t}(\theta)$ is $(y_t - x_t \beta) x_t = u_t
  x_t$; the orthogonality condition is interpretable as the
  requirement that the regressors should be uncorrelated with the
  disturbances;
\item $W$ can be any symmetric positive definite matrix, since
  the number of parameters equals the number of orthogonality
  conditions. Let's say we choose $I$.
\item The function $F(\theta, W)$ is in this case
  \[
    F(\theta, W) = \left[ \frac{1}{T} \sum_{t=1}^T (\hat{u}_t x_t) \right]^2
  \]
  and it is easy to see why OLS and GMM coincide here: the GMM
  objective function has the same minimizer as the objective function
  of OLS, the residual sum of squares. Note, however, that the two
  functions are not equal to one another: at the minimum, $F(\theta,
  W) = 0$ while the minimized sum of squared residuals is zero only in
  the special case of a perfect linear fit.
\end{itemize}

The code snippet below uses gretl's \texttt{gmm} command to make the
above operational.  The series \texttt{e} holds the ``residuals'' and
the series \texttt{x} holds the regressor.  If \texttt{x} had been a
list (or a matrix), the \texttt{orthog} statement would have generated
one orthogonality condition for each element (or column) of
\texttt{x}.
%
\begin{code}
/* initialize stuff */
series e = 0
scalar beta = 0
matrix W = I(1)

/* proceed with estimation */
gmm 
  series e = y - x*beta
  orthog e ; x
  weights W
  params beta
end gmm
\end{code}


\section{TSLS as GMM}
\label{sec:gmm-tsls}

Moving closer to the proper domain of GMM, we now consider two-stage
least squares (TSLS) as a case of GMM.  

TSLS is employed in the case where one wishes to estimate a linear
model of the form $y_t = X_t \beta + u_t$, but where one or more of
the variables in the matrix $X$ are potentially
endogenous---correlated with the error term, $u$.  We proceed by
identifying a set of instruments, $Z_t$, which are explanatory for the
endogenous variables in $X$ but which are plausibly uncorrelated with
$u$.  The classic two-stage procedure is (1) regress the endogenous
elements of $X$ on $Z$; then (2) estimate the equation of interest,
with the endogenous elements of $X$ replaced by their fitted values
from (1).

An alternative perspective is given by GMM.  We define the residual
$\hat{u}_t$ as $y_t - X_t \hat{\beta}$, as usual.  But instead of
relying on $E(u|X) = 0$ as in OLS, we base estimation on the condition
$E(u|Z) = 0$.  In this case it is natural to base the initial
weighting matrix on the covariance matrix of the instruments.
Listing~\ref{gmm-tsls-ex} presents a model from Stock and Watson's
\textit{Introduction to Econometrics}.  The demand for cigarettes is
modeled as a linear function of the logs of price and income; income
is treated as exogenous while price is taken to be endogenous and two
measures of tax are used as instruments.  Since we have two
instruments and one endogenous variable the model is over-identified.

In the GMM context, this happens when you have more orthogonality
conditions than parameters to estimate. If so, asymptotic efficiency
gains can be expected by iterating the procedure once or more. This is
accomplished by specifying, after the \cmd{end gmm} statement, two
mutually exclusive options: \option{two-step} or \option{iterate},
whose meaning should be obvious.  Note that when the problem is
over-identified, the weights matrix will influence the solution you
get from the 1- and 2-step procedures.

\tip{In cases other than one-step estimation the specified weights
  matrix will be overwritten with the \emph{final} weights on
  completion of the \texttt{gmm} command. If you wish to execute more
  than one GMM block with a common starting-point it is therefore
  necessary to reinitialize the weights matrix between runs.}


Partial output from this script is shown in~\ref{gmm-tsls-out}.  The
estimated standard errors from GMM are robust by default; if we supply
the \option{robust} option to the \texttt{tsls} command we get
identical results.\footnote{The data file used in this example is
  available in the Stock and Watson package for gretl.  See
  \url{http://gretl.sourceforge.net/gretl_data.html}.}

After the \texttt{end gmm} statement two mutually
exclusive options can be specified: \option{two-step} or
\option{iterate}, whose meaning should be obvious.

\begin{script}[htbp]
  \scriptcaption{TSLS via GMM}
  \label{gmm-tsls-ex}
\begin{scode}
open cig_ch10.gdt
# real avg price including sales tax
ravgprs = avgprs / cpi
# real avg cig-specific tax
rtax = tax / cpi
# real average total tax
rtaxs = taxs / cpi
# real average sales tax
rtaxso = rtaxs - rtax
# logs of consumption, price, income
lpackpc = log(packpc)
lravgprs = log(ravgprs)
perinc = income / (pop*cpi)
lperinc = log(perinc)
# restrict sample to 1995 observations
smpl --restrict year==1995
# Equation (10.16) by tsls
list xlist = const lravgprs lperinc
list zlist = const rtaxso rtax lperinc
tsls lpackpc xlist ; zlist --robust

# setup for gmm
matrix Z = { zlist }
matrix W = inv(Z'Z)
series e = 0
scalar b0 = 1
scalar b1 = 1
scalar b2 = 1

gmm e = lpackpc - b0 - b1*lravgprs - b2*lperinc
  orthog e ; Z
  weights W
  params b0 b1 b2
end gmm 
\end{scode}
\end{script}

\begin{script}[htbp]
  \caption{TSLS via GMM: partial output}
  \label{gmm-tsls-out}
\begin{code}
Model 1: TSLS estimates using the 48 observations 1-48
Dependent variable: lpackpc
Instruments: rtaxso rtax 
Heteroskedasticity-robust standard errors, variant HC0

      VARIABLE       COEFFICIENT        STDERROR      T STAT   P-VALUE

  const                 9.89496          0.928758     10.654  <0.00001 ***
  lravgprs             -1.27742          0.241684     -5.286  <0.00001 ***
  lperinc               0.280405         0.245828      1.141   0.25401

Model 2: 1-step GMM estimates using the 48 observations 1-48
e = lpackpc - b0 - b1*lravgprs - b2*lperinc

      PARAMETER       ESTIMATE          STDERROR      T STAT   P-VALUE

  b0                    9.89496          0.928758     10.654  <0.00001 ***
  b1                   -1.27742          0.241684     -5.286  <0.00001 ***
  b2                    0.280405         0.245828      1.141   0.25401

  GMM criterion = 0.0110046
\end{code}
\end{script}


\section{Covariance matrix options}
\label{sec:gmm-vcv}

The covariance matrix of the estimated parameters depends on the
choice of $W$ through
\begin{equation}
  \label{eq:gmmest-vcv}
    \hat{\Sigma} = (J'WJ)^{-1} J'W\Omega W J (J'WJ)^{-1}
\end{equation}
where $J$ is a Jacobian term
\[
  J_{ij} = \pder{\bar{f}_i}{\theta_j}
\]
and $\Omega$ is the long-run covariance matrix of the orthogonality
conditions. 

Gretl computes $J$ by numeric differentiation (there is no
provision for specifying a user-supplied analytical expression for $J$
at the moment). As for $\Omega$, a consistent estimate is needed. The
simplest choice is the sample covariance matrix of the $f_t$s:
\begin{equation}
  \label{eq:gmm-hcvar}
    \hat{\Omega}_0(\theta) = \frac{1}{T} \sum_{t=1}^T f_t(\theta) f_t(\theta)'
\end{equation}

This estimator is robust with respect to heteroskedasticity, but not
with respect to autocorrelation.  A heteroskedasticity- and
autocorrelation-consistent (HAC) variant can be obtained using the
Bartlett kernel or similar.  A univariate version of this is used in
the context of the \texttt{lrvar()} function---see equation
(\ref{eq:scalar-lrvar}).  The multivariate version is set out in
equation (\ref{eq:gmm-hacvar}).

\begin{equation}
  \label{eq:gmm-hacvar}
    \hat{\Omega}_k(\theta) = \frac{1}{T} 
    \sum_{t=k}^{T-k} \left[ \sum_{i=-k}^k w_i f_t(\theta) f_{t-i}(\theta)'  \right] ,
\end{equation}

Gretl computes the HAC covariance matrix by default when a GMM
model is estimated on time series data.  You can control the kernel
and the bandwidth (that is, the value of $k$ in \ref{eq:gmm-hacvar})
using the \texttt{set} command.  See chapter~\ref{chap:robust_vcv} for
further discussion of HAC estimation.  You can also ask gretl
\emph{not} to use the HAC version by saying
%
\begin{code}
set force_hc on
\end{code}

\section{A real example: the Consumption Based Asset Pricing Model}
\label{sec:gmm-CBAPM}

To illustrate gretl's implementation of GMM, we will replicate
the example given in chapter 3 of \cite{hall05}. The model to estimate
is a classic application of GMM, and provides an example of a case
when orthogonality conditions do not stem from statistical
considerations, but rather from economic theory.

A rational individual who must allocate his income between consumption
and investment in a financial asset must in fact choose the consumption
path of his whole lifetime, since investment translates into future
consumption. It can be shown that an optimal consumption path should
satisfy the following condition:
\begin{equation}
  \label{eq:gmm-CBAPM}
  p U'(c_t) = \delta^k E\left[ r_{t+k} U'(c_{t+k}) | \mathcal{F}_t
  \right] ,
\end{equation}
where $p$ is the asset price, $U(\cdot)$ is the individual's utility
function, $\delta$ is the individual's subjective discount rate and
$r_{t+k}$ is the asset's rate of return between time $t$ and time
$t+k$. $\mathcal{F}_t$ is the \emph{information set} at time $t$;
equation (\ref{eq:gmm-CBAPM}) says that the utility ``lost'' at time
$t$ by purchasing the asset instead of consumption goods must be
matched by a corresponding increase in the (discounted) future utility
of the consumption financed by the asset's return. Since the future is
uncertain, the individual considers his expectation, conditional on
what is known at the time when the choice is made.

We have said nothing about the nature of the asset, so equation
(\ref{eq:gmm-CBAPM}) should hold whatever asset we consider; hence, it
is possible to build a system of equations like (\ref{eq:gmm-CBAPM})
for each asset whose price we observe.

If we are willing to believe that
\begin{itemize}
\item the economy as a whole can be represented as a single gigantic
  and immortal representative individual, and
\item the function $U(x) = \frac{x^{\alpha} - 1 }{\alpha}$ is a
  faithful representation of the individual's preferences,
\end{itemize}
then, setting $k=1$, equation (\ref{eq:gmm-CBAPM}) implies the
following for any asset $j$:
\begin{equation}
  \label{eq:gmm-CBAPM-est}
  E\left[ \delta \frac{r_{j,t+1}}{p_{j,t}} \left(\frac{C_{t+1}}{C_{t}}
    \right)^{\alpha - 1} \bigg| \mathcal{F}_t \right] = 1 ,
\end{equation}
where $C_t$ is aggregate consumption and $\alpha$ and $\delta$ are the
risk aversion and discount rate of the representative individual. In
this case, it is easy to see that the ``deep'' parameters $\alpha$ and
$\delta$ can be estimated via GMM by using
\[
  e_t = \delta \frac{r_{j,t+1}}{p_{j,t}} \left(\frac{C_{t+1}}{C_{t}}
    \right)^{\alpha - 1} - 1
\]
as the moment condition, while any variable known at time $t$ may serve as
an instrument.


\begin{script}[htbp]
  \scriptcaption{Estimation of the Consumption Based Asset Pricing Model}
  \label{gmm-CBAPM-script}
\begin{scode}
open hall.gdt
set force_hc on

scalar alpha = 0.5
scalar delta = 0.5
series e = 0

list inst = const consrat(-1) consrat(-2) ewr(-1) ewr(-2)

matrix V0 = 100000*I(nelem(inst))
matrix Z = { inst }
matrix V1 = $nobs*inv(Z'Z)

gmm e = delta*ewr*consrat^(alpha-1) - 1
  orthog e ; inst
  weights V0
  params alpha delta
end gmm

gmm e = delta*ewr*consrat^(alpha-1) - 1
  orthog e ; inst
  weights V1
  params alpha delta
end gmm

gmm e = delta*ewr*consrat^(alpha-1) - 1
  orthog e ; inst
  weights V0
  params alpha delta
end gmm --iterate

gmm e = delta*ewr*consrat^(alpha-1) - 1
  orthog e ; inst
  weights V1
  params alpha delta
end gmm --iterate
\end{scode}
\end{script}

\begin{script}[htbp]
  \caption{Estimation of the Consumption Based Asset Pricing Model --
  output}
  \label{gmm-CBAPM-out}
\begin{outbit}
Model 1: 1-step GMM estimates using the 465 observations 1959:04-1997:12
e = d*ewr*consrat^(alpha-1) - 1

      PARAMETER       ESTIMATE          STDERROR      T STAT   P-VALUE

  alpha                -3.14475          6.84439      -0.459   0.64590
  d                     0.999215         0.0121044    82.549  <0.00001 ***

  GMM criterion = 2778.08

Model 2: 1-step GMM estimates using the 465 observations 1959:04-1997:12
e = d*ewr*consrat^(alpha-1) - 1

      PARAMETER       ESTIMATE          STDERROR      T STAT   P-VALUE

  alpha                 0.398194         2.26359       0.176   0.86036
  d                     0.993180         0.00439367  226.048  <0.00001 ***

  GMM criterion = 14.247

Model 3: Iterated GMM estimates using the 465 observations 1959:04-1997:12
e = d*ewr*consrat^(alpha-1) - 1

      PARAMETER       ESTIMATE          STDERROR      T STAT   P-VALUE

  alpha                -0.344325         2.21458      -0.155   0.87644
  d                     0.991566         0.00423620  234.070  <0.00001 ***

  GMM criterion = 5491.78
  J test: Chi-square(3) = 11.8103 (p-value 0.0081)

Model 4: Iterated GMM estimates using the 465 observations 1959:04-1997:12
e = d*ewr*consrat^(alpha-1) - 1

      PARAMETER       ESTIMATE          STDERROR      T STAT   P-VALUE

  alpha                -0.344315         2.21359      -0.156   0.87639
  d                     0.991566         0.00423469  234.153  <0.00001 ***

  GMM criterion = 5491.78
  J test: Chi-square(3) = 11.8103 (p-value 0.0081)
\end{outbit}
\end{script}

In the example code given in \ref{gmm-CBAPM-script}, we replicate
selected portions of table 3.7 in \cite{hall05}.  The variable
\texttt{consrat} is defined as the ratio of monthly consecutive real
per capita consumption (services and nondurables) for the US, and
\texttt{ewr} is the return--price ratio of a fictitious asset
constructed by averaging all the stocks in the NYSE.  The instrument
set contains the constant and two lags of each variable.

The command \texttt{set force\_hc on} on the second line of the script
has the sole purpose of replicating the given example: as mentioned
above, it forces gretl to compute the long-run variance of the
orthogonality conditions according to equation (\ref{eq:gmm-hcvar})
rather than (\ref{eq:gmm-hacvar}).

We run \texttt{gmm} four times: one-step estimation for each of two
initial weights matrices, then iterative estimation starting from each
set of initial weights.  Since the number of orthogonality conditions
(5) is greater than the number of estimated parameters (2), the choice
of initial weights should make a difference, and indeed we see fairly
substantial differences between the one-step estimates (Models 1 and
2).  On the other hand, iteration reduces these differences almost to
the vanishing point (Models 3 and 4). 

Part of the output is given in \ref{gmm-CBAPM-out}.  It should be
noted that the $J$ test leads to a rejection of the hypothesis of
correct specification.  This is perhaps not surprising given the
heroic assumptions required to move from the microeconomic principle
in equation (\ref{eq:gmm-CBAPM}) to the aggregate system that is
actually estimated.


\section{Caveats}
\label{sec:gmm-caveat}

A few words of warning are in order: despite its ingenuity, GMM is
possibly the most fragile estimation method in econometrics. The
number of non-obvious choices one has to make when using GMM is large,
and in finite samples each of these can have dramatic consequences for
the eventual output. Some of the factors that may affect the results
are:
\begin{enumerate}
\item Orthogonality conditions can be written in more than one way:
  for example, if $E(x_t - \mu) = 0$, then $E(x_t/\mu - 1) =
  0$ holds too. It is possible that a different specification of the
  moment conditions leads to different results.
\item As with all other numerical optimization algorithms, weird
  things may happen when the objective function is nearly flat in some
  directions or has multiple minima. BFGS is usually quite good, but
  there is no guarantee that it always delivers a sensible solution,
  if one at all.
\item The 1-step and, to a lesser extent, the 2-step estimators may be
  sensitive to apparently trivial details, like the re-scaling of the
  instruments. Different choices for the initial weights matrix can
  also have noticeable consequences.
\item With time-series data, there is no hard rule on the appropriate
  number of lags to use when computing the long-run covariance matrix
  (see section \ref{sec:gmm-vcv}). Our advice is to go by trial and
  error, since results may be greatly influenced by a poor choice.
\end{enumerate}

One of the consequences of this state of things is that replicating
well-known published studies may be extremely difficult. Any
non-trivial result is virtually impossible to reproduce unless all
details of the estimation procedure are carefully recorded.

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