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3.? ECONOMETRIC METHOD
In this following section we estimate models of relative regional log earnings (taken from the
New Earnings Survey, see Data Appendix for details, and Cameron and Muellbauer (2000) for
discussion of regional data problems) for two types of workers and relative regional unemployment
rates using data for 1972 to 1995 for the ten regions of
general forms:
Thus, the long-run solution has the form:
Δyit=α(θi+∑ki=1βjxjit-1-yit-1)+∑ms=1γsΔyit-s+∑ms=1∑kj=1γjsΔxjit-s+εit (1)
?Thus, the long-run solution has the form
???????????????? yit=θi+∑kj=1βjxjit+ηit ??(2 )
where the three y variables are, respectively, relative log-earnings for full-time men and womenly
and the unemployment rate in each region minus the rate in
disturbance terms,εit , are distributed independently across groups and time and that ηit is a
stationary process (see Pesaran, Shin and Smith, 1999). For each y variable, the x variables
include the relevant complementary elements of the other two y variables; relative house prices
weighted by lagged owner-occupation in the
the unemployment equation; the relative proportion of employment in the production sector and its
interaction with the log real exchange rate; the relative proportion of employment in banking and
financial services and its interactions with bank base rate, a proxy for financial liberalization, and
real
proportion of manual workers. The x variables also include interactions between the relative
mortgage debt to earnings ratio with the average mortgage interest rate. All the x variables test as
being I(1) in levels and I(0) in differences using the panel unit root test suggested by Im, Pesaran
and Shin (1997), as do the y variables, (see the Data Appendix).
其中的一小段
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4.1.two Models of Relative Regional Earnings
Full-time Men's Relative Earnings
The men's earnings version of equation (1) was reduced to the more parsimonious specification
reported in Table I, column 1. There is a significant equilibrium-correction term with a around 0.45
which implies that earnings return quickly to their equilibrium levels. There are two points to note
here. First, the presence of this term means that the equation can be interpreted as a conditional
convergence regression, with the implication that relative steady-state income is determined by
the other variables in the model. Second, this is evidence against Blanchflower and Oswald's
contention that autoregression is unimportant in wage equations (1994, p. 284).
? The lagged log level of the deviation of regional unemployment from the national level has a
significant and negative effect with a long-run coefficient around -0.034, one third of the一0.1
figure claimed by Blanchflower and Oswald as a robust order of magnitude of the slope of the wage
curve. Contemporaneous changes in unemployment were instrumented using the forecast changes
from the model described in the next section, but were found to be insignificant. One possible
reason for this is that the NES earnings data are observed in April while the unemployment rates
are annual averages. However, there is a strong negative effect from the change in the previous
year's relative unemployment rate.
? Turning to the housing market effects, lagged relative house prices have a positive and highly
significant effect on relative earnings, and this effect has become stronger as the proportion of
owner-occupation in the
produces a more significant coefficient here than if the proportion of owner-occupation in the
region is used or than if relative log house prices are unweighted. At a
proportion of 0.68, the long-run effect of relative log house prices on men's relative full-time
earnings is 0.075. Relative mortgage costs in the previous year have a positive effect on relative
earnings. But a rise in mortgage interest rates 2 years earlier has a temporary negative effect
on relative earnings in regions with high ratios of mortgage debt to earnings. This appears to
reflect the long-term recessionary implications of higher interest rates on regions with high debt
to income ratios. Last year's expectation of a rise in this year's bank base rate, interacted with
the relative proportion of employment in banking and financial services, has a negative effect
on relative earnings. There is no evidence of a spill-over effect from women's relative regional
earnings.
? Lastly, we have the composition effects. Since this is a model of men's earnings, it is interesting
that regions with high proportions of part-time women have lower relative men's earnings,
suggesting an element of substitution between these groups of workers, or that a high proportion
of part-time women signals weak labour demand. We also find that men's earnings in regions with
more production workers suffer more when competitiveness falls (that is, the log real exchange
rate rises) and that men's earnings in regions with more banking and financial sector workers do
better when there is financial liberalization.
? Column 2 of Table I presents the short-sample estimate of the model, from 1972 to 1987. This
omits the peak of the 1980s house price boom and the 1990s housing market crisis, a severe test
of parameter stability. There are no significant differences in estimated coefficients, though the
point estimate of the lagged weighted change in mortgage interest rates falls, and the restriction
of no structural break cannot be rejected at the 5% level, Fso,iss=1.10 [P=0.31].
14 6L2008/6/22 0:58:00
LZ，这个我懂不懂好难讲，我的确看懂了，但是你这个有涉及到equillibirum correction model吧？就是那个均衡修正模型，我没学过。我们QA没讲到那么深。如果LZ只是想要翻译我倒是可以尽一份力
邮箱反白 kyukoryu@hotmail.com
我撑不住了，我这三点了……先下去睡了……
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