Department of Economics, Norwegian University of Science and Technology, Trondheim, Norway

Department of Community Dentistry, University of Oslo, Oslo, Norway

Department of Obstetrics and Gynecology, Institute of Clinical Medicine, Akershus University Hospital, Lørenskog, Norway

Department of Mental Health, Norwegian Institute of Public Health, Oslo, Norway

Abstract

Background

There has been a considerable decline in fetal and neonatal mortality in the Western world. The authors hypothesized that this decline has been largest for boys, since boys have a higher risk of fetal and neonatal death.

Methods

The authors used data from the Medical Birth Registry about all births in Norway to study changes during 1967–2005 in mortality for boys and girls from the 23rd week of pregnancy until one month after birth. Absolute and relative yearly changes in fetal and neonatal death rates were estimated separately for boys and girls.

Results

From 1967 to 2005, the average annual reduction in the overall death rate was greater for boys: 0.47 per 1000 boys (95% CI: 0.45, 0.48) and 0.37 per 1000 girls (95% CI: 0.35, 0.39). These estimates were not affected by adjustments made for changes over time in maternal characteristics. The convergence in death rates by sex was strongest for the first week after birth: average annual reduction in the early neonatal death rate was 0.24 per 1000 boys (95% CI: 0.23, 0.25) and 0.17 per 1000 girls (95% CI: 0.16, 0.18). The death rates for boys and girls also converged during pregnancy and from one week to one month after birth. The relative reduction in death rates was quite similar for boys and girls: the overall death rate fell annually by 4.4% (95% CI: 4.3, 4.6%) for boys and by 4.2% (95% CI: 4.0, 4.4%) for girls.

Conclusions

During the period 1967–2005, the absolute reduction in fetal and neonatal death rates was greatest for boys. The relative reduction in mortality was about the same for both sexes, but the absolute reduction was greatest for boys since the mortality for boys began at a higher level. The convergence of death rates was not due to changes in the composition of mothers, suggesting that convergence has been caused by technological progress.

Background

During the last decades, there has been a considerable decline in fetal and neonatal mortality in the Western world

Among all deliveries by Norwegian women during the period 1967–2005, we studied whether the decline in fetal and neonatal death rates differed between boys and girls. Separate analyses were performed for pre-term births, term births, the early neonatal period and the late neonatal period.

Methods

Design

We performed a population-based follow-up study by utilizing data from two national Norwegian registers: the Medical Birth Registry of Norway

Study population

Since 1967, all births after 16 weeks of gestation have been reported to the Medical Birth Registry of Norway. Our study population comprised all births from 23 weeks of gestation during the period 1967 to 2005, a total of 2,263,736 births. We excluded all mothers that immigrated to Norway (N = 144,858). We want to focus on trends in long term mortality caused by improvements in maternity care and infant care. From 1967 to 2005, the proportion of deliveries by immigrant mothers increased from 2.3% to 15.6%. Children of immigrant mothers have, on average, higher mortality than the rest of the population, and there has been considerable variation over time in the countries that immigrants to Norway come from

Pregnancies recorded as lasting more than 43 weeks were excluded (N = 27,341), since gestational length may have been miscoded in some of these pregnancies. We could not with certainty determine which were miscoded, so we excluded all these pregnancies. Also, we omitted births with unknown sex of the child (N = 260) and without information on gestational age at birth (N = 116,466) leaving 1,974,811 births for the analysis.

Study factors

Gestational age at birth was calculated from the date of the last menstrual period for births during the years 1967 to 1998. For the years 1999 to 2005, gestational age at birth was based on estimates of term date at routine fetal ultrasonographical examinations in weeks 17–19 of pregnancy. Information from ultrasonographical examinations was available for 97.6% of the women, and for the remaining women the date of the last menstrual period was used. Information on vital status at birth was obtained from the Medical Birth Registry. Information about infant death and the age of the infant at death were obtained from the Central Person Registry.

We adjusted for maternal age, parity, plurality and maternal education at delivery, since these factors are associated with offspring death

Statistical methods

We carried out separate analyses for each of the following periods of pregnancy: 23–28 weeks, 29–36 weeks and 37–43 weeks. In each analysis, only fetuses that were alive in utero at the beginning of the period were included. The outcome variable distinguished between pregnancies where the fetus died in utero or during birth and pregnancies where the offspring was born alive or was still alive in utero by the end of each period of pregnancy. For infants who were born alive, we did separate analyses for the following time periods after birth: ≤ 1 week and 1 week-1 month. For these analyses the outcome variable distinguished between infants who died during the period and infants who were alive at the end of the period.

To estimate absolute yearly change in fetal and neonatal death rates, the following linear probability model

where Mortality_{it} is a dummy variable that is one if fetus/infant _{it} (Girl_{it}) is a dummy variable turned on if the fetus/infant was a boy (girl), ϵ_{it} is an error term, and α_{1} – α_{6} are parameters to be estimated. α_{1} can be interpreted as the predicted death rate for boys in 1967, that is, the death rate that would have prevailed if the error term had been zero. α_{2} is the corresponding predicted death rate for girls in 1967. The predicted death rates in year t are α_{1} + (t-1967) α_{3} + (t-1967)^{2} α_{5} for boys and α_{2} + (t-1967) α_{4} + (t-1967)^{2} α_{6} for girls, implying that predicted death rates fell by 38 α_{3} + 38^{2} α_{5} for boys and 38 α_{4} + 38^{2} α _{6} for girls during 1967–2005. The absolute yearly change in death rates in year t is α_{3} + 2 (t-1967) α_{5} for boys and α_{4} + 2 (t-1967) α_{6} for girls. α_{3} and α_{4} can be interpreted as absolute yearly changes in death rates in 1967. In 2005, absolute yearly changes in death rates were α_{3} + 76 α_{5} for boys and α_{4} + 76 α_{6} for girls.

We also estimated a simplified linear probability model where α_{5} and α_{6} are set equal to zero. In the simplified specification, absolute yearly changes in death rates are constant and equal to α_{3} for boys and α_{4} for girls. The estimates of α_{3} and α_{4} in this model can therefore be interpreted as average yearly changes in absolute death rates during the whole period from 1967 to 2005.

Since there is considerable random variation between years in death rates, we report predicted rather than actual death rates for individual years. Adjustment for the effects of maternal age, parity, plurality and maternal education on absolute yearly changes in death rates was made by including these variables as covariates in the linear probability models.

To obtain estimates of relative yearly change in death rates, the following logistic regression model was fitted:

where Λ is the logistic cumulative distribution function _{2})-1 can be interpreted as predicted excess death rate for boys in 1967, whereas exp(α_{3})-1 and exp(α_{4})-1 are the relative yearly change in death rates for boys and girls respectively. Adjusted estimates of relative yearly changes were obtained by including maternal factors as covariates in the logistic regression model.

We estimated the sex-specific number of fetal and infant deaths prevented from 1967 to 2005 by comparing two scenarios: one scenario where, for each year from 1967–2005, mortality was set equal to predicted mortality for 1967, and another scenario where mortality was set equal to predicted mortality in the actual year. In these calculations, the simplified linear probability model was used.

The study was approved by the Regional Committee for Medical and Health Research Ethics (Reference number 603-07276a 1.2007.2366).

Results

During the study period 1967 to 2005, fetal mortality was higher for boys than for girls in pre-term pregnancies (Table

**Up to birth**

**After birth**

**23 - 28 weeks**

**29-36 weeks**

**≥ 37 weeks**

**≤ 1 week**

**1 week - 1 month**

**Boys**

**Girls**

**Boys**

**Girls**

**Boys**

**Girls**

**Boys**

**Girls**

**Boys**

**Girls**

Number at risk

1,015,083

959,728

1,014,279

959,035

1,012,028

957,188

1,009,708

954,987

1,004,798

951,578

Number of deaths

804

693

2,251

1,847

2,320

2,201

4,910

3,409

959

637

Percentage deaths

0.08

0.07

0.22

0.19

0.23

0.23

0.49

0.36

0.10

0.07

Overall decline in mortality

Table

**Absolute change**

**Relative change**

**Coefficient**

**95% CI**

**Coefficient**

**95% CI**

**Coefficient**

**95% CI**

Linear probability and logistic regression analyses. Pregnancies with gestational length >22.

Boy

24.094

23.567, 24.622

19.647

19.272, 20.002

1.245

1.191, 1.301

Girl

19.406

18.863, 19.949

15.958

15.583, 16.334

1.000

Boy * (Year - 1967)

−1.116

−1.183, -1.048

−0.466

−0.483,−0.449

0.956

0.954, 0.957

Girl * (Year-1967)

−0.867

−0.936, −0.798

−0.370

−0.387, −0.353

0.958

0.956, 0.960

Boy * (Year - 1967)^{2}

0.017

0.015, 0.019

Girl * (Year - 1967)^{2}

0.013

0.011, 0.015

Test for equal average trends for boys and girls:

p < 0.001

p = 0.132

N

1,974,811

Estimated time trends of overall death rates for boys and girls

**Estimated time trends of overall death rates for boys and girls.**

The second column in Table

The annual relative change in the death rate was about the same for boys and girls (the third column of Table

When the maternal factors, age, parity, plurality and education at delivery, were included as covariates, the estimated annual reductions in absolute and relative death rates changed very little (Table

**Absolute change**

**Relative change**

**Coefficient**

**95% CI**

**Coefficient**

**95% CI**

Linear probability and logistic regression analyses. Adjustment made for maternal age (five-year intervals), parity (first child, other children), plurality (single, multiple births) and maternal education (compulsory school only, upper secondary education, university/college education, unknown education).

Boy

19.647

19.052, 20.242

1.242

1.188, 1.298

Girl

15.958

15.356, 16.560

1.000

Boy * (Year - 1967)

−0.460

−0.477, −0.442

0.958

0.956, 0.960

Girl * (Year - 1967)

−0.365

−0.383, −0.357

0.960

0.958, 0.962

Test for equal average trends for boys and girls:

p < 0.001

p = 0.132

N

1,974,811

Decline in fetal mortality

In 1967, boys had the highest fetal mortality. The predicted death rates in 1967 were (all per thousand): 1.1 for boys (95% CI: 1.0, 1.2) and 1.0 for girls (95% CI: 0.9, 1.1) in weeks 23–28, 3.8 for boys (95% CI: 3.7, 4.0) and 3.4 for girls (95% CI: 3.3, 3.6) in weeks 29–36, and 4.1 for boys (95% CI: 3.9, 4.3) and 3.9 for girls (95% CI: 3.8, 4.1) in weeks 37–43 (Table

**23-28 weeks**

**29-36 weeks**

**≥ 37 weeks**

**Absolute change**

**Relative change**

**Absolute change**

**Relative change**

**Absolute change**

**Relative change**

**Coefficient**

**95% CI**

**Coefficient**

**95% CI**

**Coefficient**

**95% CI**

**Coefficient**

**95% CI**

**Coefficient**

**95% CI**

**Coefficient**

**95% CI**

Non-immigrants in Norway, 1967–2005. Linear probability and logistic regression analyses. Pregnancies with gestational length >22 weeks.

Boy

1.146

1.046, 1.247

1.184

0.997, 1.407

3.847

3.680, 4.013

1.102

1.000, 1.214

4.105

3.930, 4.280

1.062

0.969, 1.164

Girl

0.982

0.879, 1.086

1.000

3.441

3.270, 3.612

1.000

3.943

3.763, 4.123

1.000

Boy * (Year - 1967)

−0.019

−0.024, −0.015

0.975

0.969, 0.981

−0.089

−0.096, −0.081

0.959

0.955, 0.962

−0.099

−0.107, −0.091

0.955

0.951, 0.959

Girl * (Year - 1967)

−0.014

−0.019, −0.009

0.980

0.974, 0.987

−0.082

−0.090, −0.075

0.955

0.951, 0.959

−0.089

−0.098, −0.081

0.960

0.956, 0.963

Test for equal average trends for boys and girls:

p = 0.131

p = 0.142

p = 0.274

p = 0.242

p = 0.119

p = 0.081

N

1,974,811

1,973,314

1,969,216

The annual reduction in death rates during 1967–2005 was 0.019 per 1000 boys (95% CI: 0.015, 0.024) and 0.014 per 1000 girls (95% CI: 0.009, 0.019) in weeks 23–28, 0.089 per 1000 boys (95% CI: 0.081, 0.096) and 0.082 per 1000 girls (95% CI: 0.075, 0.090) in weeks 29–36 and 0.099 per 1000 boys (95% CI: 0.091, 0.107) and 0.089 per 1000 girls (95% CI: 0.081, 0.098) in weeks 37–43 (Table

Decline in early neonatal mortality

In 1967, the predicted death rates during the first week after birth were 9.3 of 1000 new-born boys (95% CI: 9.1, 9.5) and 6.7 of 1000 new-born girls (95% CI: 6.5, 6.9) (Table

**≤ 1 week**

**1 week - 1 month**

**Absolute change**

**Relative change**

**Absolute change**

**Relative change**

**Coefficient**

**95% CI**

**Coefficient**

**95% CI**

**Coefficient**

**95% CI**

**Coefficient**

**95% CI**

Linear probability and logistic regression analyses. Pregnancies with gestational length > 22 weeks.

Boy

9.305

9.067, 9.542

1.406

1.314, 1.504

1.360

1.255, 1.465

1.393

1.176, 1.649

Girl

6.705

6.460, 6.949

1.000

0.972

0.864, 1.080

1.000

Boy * (Year - 1967)

−0.241

−0.252, −0.230

0.947

0.944, 0.950

−0.022

−0.027, −0.017

0.977

0.971, 0.982

Girl * (Year - 1967)

−0.170

−0.182, −0.159

0.949

0.946, 0.952

−0.016

−0.021, −0.011

0.975

0.969, 0.982

Test for equal average trends for boys and girls:

p < 0.001

p = 0.253

p = 0.115

p = 0.729

N

1,964,695

1,956,376

Decline in late neonatal mortality

In 1967, the predicted late neonatal death rate was 1.4 per 1000 boys (95% CI: 1.3, 1.5), and 1.0 per 1000 girls (95% CI: 0.9, 1.1) (Table

Prevented numbers of fetal and neonatal deaths

Based on the results for absolute changes in death rates (Tables

**Up to birth**

**After birth**

**Total**

**23-28 weeks**

**29-36 weeks**

**≥ 37 weeks**

**≤ 1 week**

**1 week - 1 month**

Based on estimates reported in Tables

Boys

386

1,767

1,956

4,767

425

9,301

Girls

269

1,556

1,679

3,187

302

6,993

Difference

117

211

277

1,580

123

2,308

Hence, the total number of prevented deaths was 9,301 for boys and 6,993 for girls, giving a sex difference in prevented deaths of 2,308. Of all prevented deaths, 68% of the sex difference (1,580/2,308) could be explained by the greater reduction in early neonatal mortality for boys, 26% (605/2,308) could be explained by the greater reduction in fetal mortality for boys and 5% (123/2,308) by the greater reduction in late neonatal mortality for boys.

Discussion

During the period 1967–2005, the fetal and neonatal death rates in Norway fell more for boys than for girls. Thus, since 1967 the difference in mortality by sex has been reduced, and higher number of deaths among boys than girls has been prevented. The relative reduction in death rates during the period was, however, about the same for boys and girls.

Strengths/limitations

The strength of this study is the population design with information about mortality for all births in Norway during almost 40 years. We had information up until 2005. The lack of data for recent years is due to the time-consuming process of updating national registries.

We lack data on the length of gestation at birth for about five per cent of pregnancies. Both mortality and the proportion of boys are somewhat higher for these pregnancies than for the pregnancies included in our sample. The exclusion of pregnancies of unknown length could have biased the estimates of trends in excess fetal mortality for boys. However, we were able to study how exclusion of these pregnancies influenced trends in excess neonatal mortality for boys. We repeated the analysis of neonatal mortality with the extra pregnancies included, and the results turned out to be very similar to the results presented in Table

Other studies

We are not aware of other research in which sex-specific trends in fetal mortality and neonatal mortality have been estimated and compared.

Interpretation

During the last decades, there has been a large decline in fetal and neonatal mortality

There is widespread concern that technological developments that contribute to saving more infants have negative side-effects, as they may lead to an increase in the number of new-born infants with health problems

Conclusions

During the period 1967–2005, the reduction in fetal mortality and neonatal mortality was greatest for boys. The main reason for this is that mortality for boys began at a higher level. The relative reduction in mortality was about the same for both sexes.

Competing interests

The authors declare that they have no competing interests.

Authors’ contributions

FC: Conception and design of the study, analysis of the data and writing of the manuscript. JG: Conception and design of the study, acquisition of data and interpretation of analysis. AE: Contributed to conception and design of the study, interpretation of analysis and writing of the manuscript. All three authors read and approved this version of the manuscript.

Acknowledgements

We wish to thank the Medical Birth Registry of Norway and Statistics Norway for providing data and Irene Skau for technical assistance. This study had financial support from the South-Eastern Norway Health Authority; research grant number 2709002.

Pre-publication history

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