First evaluations of the monthly data for infant mortality rates in Japan after Fukushima showed significant peaks in May and December 2011 . In addition, an analysis of the numbers of live births in Fukushima Prefecture found a highly significant 15% decrease in December 2011, nine months after the nuclear disaster . These analyses, however, were based on preliminary data. Recently, the final data were published which made a re-evaluation of the data necessary.
The present work examines infant mortality rates in a defined study area around Fukushima. This study area was constructed by the author using official data on average cesium soil contamination levels. It consists of the seven Prefectures of Fukushima, Iwate, Miyagi, Gunma, Tochigi, Ibaraki and Chiba (see Figure 1). Infant mortality rates in the study area after the Fukushima disaster in March 2011 are compared with the expected trend of the data before Fukushima.
Monthly data on live births and infant deaths from 2002 through to 2012, are available at http://www.e-stat.go.jp in Japanese . The data were translated and extracted as Excel files and sent to the author by Masao Fukumoto from Berlin.
After the Chernobyl nuclear disaster in April 1986, a first increase in perinatal mortality
occurred in February 1987, 9.5 months after the accident . Accordingly, a possible
increase in infant mortality rates in Japan was not expected before the end of 2011.
To test whether infant mortality rates in 2012 in the study region differ from the trend of the
data before 2012, a common logistic regression of the data in the study region and the
control area (the rest of Japan outside the study region) was carried out with individual
intercepts and a common parameter for the temporal trend of the data before March 2011.
Seasonal fluctuations occur each year in the monthly data on live births and infant deaths.
The seasonal pattern is assumed to be equal in the study and control regions. Dummy
variables indicate the 11 months February to December (in the form feb, mar, .. , dec);
January is used as the reference month. Overall, the logistic regression model requires 14
parameters. It has the following form (notation according to statistical software R):
glm (y ~ x+feb+mar+apr+june+jun+jul+aug+sep+oct+nov+dec+study, family=binomial)
The time variable, x, is defined as calendar month minus 2000 where calendar month (t) is
expressed in fractions of a year (e.g. January 2002 means t=2002+1/24). The dummy variable
“study” denotes the data of the study area.
Figure 2 shows the trend of the data from the study and control regions and their respective
trend lines; the lower panel plots the deviations of infant mortality rates from the expected
trend in units of standard deviations (standardized residuals). Almost all residuals fall within
the range of ±2 standard deviations which shows that the model fits the data well.
The highly significant peak of infant mortality in March 2011 in the study region was likely
caused by the earthquake and tsunami. In the course of the residuals after Fukushima, a
significant maximum occurs in March 2012. In the period from December 2011 to September
2012, all residuals are positive. The increase of infant mortality in this period corresponds to
60 excess infant deaths.
Alternative approach: analysis of odds ratios
The regression model can be radically simplified if the ratio of infant mortality rates in the
study region to the rates in the control area is analyzed. Then the seasonal variations, the
time trend, and the dummy “study” can be omitted in the regression model, so only one
parameter (intercept) is needed. For computational reasons, odds ratios were evaluated
instead of rate ratios. The odds are defined as p / (1 -p ) with rate p = ID / LB. Here ID is the
number of infant deaths and LB is the number of live births. When the logarithm of the odds
ratio is used as the dependent variable in the regression model, the variance (var) takes the
following simple form:
var = 1/ID0 + 1 / ( LB0 - ID0 ) + 1/ID1 + 1 / (LB1 - ID1 )
where 1 denotes the study region and 0 (zero) the control region.
The above regression showed that infant mortality rates were only increased in 2012 with a significant peak in March. To test whether this increase is significant, the excess is modeled by a bell-shaped function (lognormal distribution). Then the regression function takes the following form (nonlinear regression):
y ~ β1 + β2 * dmar11 + β3/t/exp((ln(t) - ln(β4))^2/β5)
The dependent variable is y = ln(OR), t is time, the dummy variable dmar11 indicates March 2011, and β1 through β5 are parameters.
The model fits the data well (deviance = 110.75 with 127 degrees of freedom). Table 1 shows the regression results.
Table 1: Regression results for odds ratios
An F test with (3, 127) degrees of freedom is used to test the significance of the excess term. It yields P = 0.0086, so the increase of infant mortality in 2012 is clearly significant.
Figure 3 shows the monthly odds ratios and the deviations of the odds ratios from the expected trend.
Birth deficit in December 2011
To estimate the effect on live births in December 2011, the monthly data of live births (LB) from January 2006 to December 2011 is analyzed using Poisson regression. A dummy variable ddec11 marks December2011. The regression model allows for a linear-quadratic time trend (variables x, x2) seasonal fluctuations (dummy variables feb, mar, .. , dec). Thus, the regression model has the following form (R notation):
glm (LB ~ x+feb+mar+apr+june+jun+jul+aug+sep+oct+nov+dec+x2+ddec11, family=quasipoisson)
Since live birth data usually show considerable overdispersion, an F test is used instead of a Chisquare test to determine the P values which is achieved by the option “family=quasipoisson”. The regression results are shown in Table 2.
Table 2: Birth deficit in December 2011 in the study area
The decrease of live births in December 2011 is 10.1% and is highly statistically significant (P = 5.8 E-7).
Figure 4 shows the trend of the live births, 2006 through 2012, and the standardized residuals. The drop of live births is limited to December 2011, no appreciable deviation of live births is observed in the previous (November 2011) and the following month (January 2012) which supports the hypothesis that the birth rate is caused by an increase in spontaneous early abortions in March 2011.
To check whether the drop of live births is associated with radiation exposure, the data from the seven prefectures of the study area are evaluated individually. The results are shown in Table 3.
Table 3: Birth deficit in the Prefectures of the study area
The infant mortality rate was significantly increased in the 7 prefectures around Fukushima with largest cesium soil contamination during the first three quarters of 2012. The decline in the number of live births in December 2011 is highly statistically significant.
After Chernobyl, a highly significant 17% drop of live births was observed in Belarus in January 1987, about 9 months after the accident in April 1986 (unpublished analysis by the author, see Figure 5). It was paralleled by a highly significant increase of Down syndrome in Belarus in January 1987 . A significant trisomy 21 peak was also found in West Berlin in the same month .
The decrease of live births in Japan and the trisomy 21 peaks after Chernobyl are limited to a single month. Therefore it seems unlikely that the effect can be explained exclusively by the reluctance to have children in the aftermath of the Fukushima accident; public worry would be expected to last for several months, as is the case in Belarus, see Figure 5. Immediately after fertilization, the zygote is extremely sensitive. Radiation damage to the zygote from the high initial radiation spike following the nuclear accident can trigger early spontaneous abortions which in turn manifest as a drop of live births 9 months later.
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 Körblein A. Rückgang der Geburten in Japan 9 Monate nach Fukushima. Strahlentelex 628-629 / 2013:2-3. http://www.strahlentelex.de/Stx_13_628-629_S02-03.pdf
 Körblein A, Küchenhoff H. Perinatal mortality in Germany following the Chernobyl accident. Radiat Environ Biophys. 1997 Feb;36(1):3-7.
 Zatsepin I, Verger P, Robert-Gnansia E, Gagnière B, Tirmarche M, Khmel R, Babicheva I, Lazjuk G. Down syndrome time-clustering in January 1987 in Belarus: link with the Chernobyl accident? Reprod Toxicol. 2007 Nov-Dec;24(3-4):289-95.
 Sperling K, Pelz J, Wegner RD, Dörries A, Grüters A, Mikkelsen M. Significant increase in trisomy 21 in Berlin nine months after the Chernobyl reactor accident: temporal correlation or causal relation? BMJ. 1994 Jul 16;309(6948):158-62.
Alfred Körblein February 2014
Increased infant mortality and decline in birth rate after Fukushima
Alfred Körblein December 2012