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Biodemography and Social Biology
ISSN: 1948-5565 (Print) 1948-5573 (Online) Journal homepage: http://www.tandfonline.com/loi/hsbi20
Parity Conditions the Risk for Low Birth Weight
after Maternal Exposure to Air Pollution
Anna Merklinger-Gruchala, Grazyna Jasienska & Maria Kapiszewska
To cite this article: Anna Merklinger-Gruchala, Grazyna Jasienska & Maria Kapiszewska
(2017) Parity Conditions the Risk for Low Birth Weight after Maternal Exposure to Air Pollution,
Biodemography and Social Biology, 63:1, 71-86, DOI: 10.1080/19485565.2016.1264872
To link to this article: http://dx.doi.org/10.1080/19485565.2016.1264872
Published online: 13 Mar 2017.
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Parity Conditions the Risk for Low Birth Weight after Maternal
Exposure to Air Pollution
, Grazyna Jasienska
and Maria Kapiszewska
Department of Pediatrics, Faculty of Medicine, Andrzej Frycz Modrzewski Krakow University, Krakow, Poland;
Department of Environmental Health, Faculty of Health Sciences, Jagiellonian University Medical College,
Department of Health and Medical Sciences, Andrzej Frycz Modrzewski Krakow University,
Multiparous mothers have greater umbilical blood ﬂow and thus more
eﬃcient transport of pollutants than primiparous mothers. We tested a
hypothesis that multiparous mothers are more prone to have an infant
with low birth weight (LBW) after prenatal exposure to air pollution. A
study was conducted on a representative group of more than 74,000
singleton, live, full-term infants. Birth data were obtained from the birth
registry, while pollution data were from an environmental monitoring
system (Poland). Multiple comparisons were controlled by the false
discovery rate procedure (FDR). After standardization, the harmful eﬀect
of carbon monoxide (CO) on the odds ratio (OR) for LBW was seen
among the multiparous mothers (OR = 1.28; 95% CI 1.06–1.54), while
in primiparous mothers itwas nonsigniﬁcant. The eﬀect of CO on the OR
for LBW diﬀered according to parity, which was conﬁrmed by the test for
interaction (FDR-adjusted p= 0.03). The interaction between parity and
sulfur dioxide (SO
) was statistically nonsigniﬁcant (FDR-adjusted
p= 0.08). Multiparous mothers may be more vulnerable to CO than
primiparous mothers. Parity may be the modiﬁer of the association
between pollutants and the risk of LBW.
A large body of literature has documented the association between atmospheric pollution
and human reproduction (Woodruﬀet al. 2010). In particular, female reproductive
physiology is sensitive to perceived environmental conditions, including lifestyle
(Jasienska et al. 2006; Kapiszewska et al. 2006; Merklinger-Gruchala et al. 2008;
Ziomkiewicz et al. 2008) and air pollution (Tomei et al. 2006).
Among the health indicators that can be used to evaluate the eﬀects of environmental
pollution on pregnancy outcome, birth weight is one of the most useful. It marks the end
point of intrauterine growth, and low birth weight may reﬂect fetal growth restriction,
especially if a neonate is delivered at term. It is also an indicator of infant mortality and
morbidity, and health in later life (Wells 2000).
It is well documented that birth weight of infants and the risk of low birth weight
(LBW), deﬁned as birth weight below 2,500 g, is negatively aﬀected by maternal exposure
to carbon monoxide (CO), sulfur dioxide (SO
), or particulate matters in the ambient air
CONTACT Anna Merklinger-Gruchala, Ph.D. firstname.lastname@example.org Department of Pediatrics, Faculty of
Medicine, Andrzej Frycz Modrzewski Krakow University, ul. Herlinga Grudzinskiego 1, Krakow 30-705, Poland.
BIODEMOGRAPHY AND SOCIAL BIOLOGY
2017, VOL. 63, NO. 1, 71–86
© 2017 Society for Biodemography and Social Biology
during pregnancy (Bobak and Leon 1999; Lin et al. 2004; Merklinger-Gruchala and
Kapiszewska 2015; Morello-Frosch et al. 2010; Stieb et al. 2012).
CO and SO
are among the major air pollutants. CO crosses the placenta rapidly,
inducing systematic oxidative stress (André et al. 2011), while SO
metabolites may act as
inhibitors of antioxidants throughout the pregnancy (Mohorovic and Micovic 2012). In
rats, such induced oxidative stress has been shown to trigger changes in vascular com-
pliance and endothelium-dependent vasoconstriction (Johnson and Johnson 2003).
Moreover, CO causes the rapid accumulation of carboxyhemoglobin and reduces the
oxygen-carrying capacity of fetal blood (Rudra et al. 2010). After CO intoxication fetal
concentrations of carboxyhemoglobin can be two-fold higher than maternal levels
(Aubard and Magne 2000). Therefore, pollution-induced placental dysfunction and
changes in the eﬃciency of the oxygen-carrying capacity of uteroplacental blood ﬂow
may prevent the fetus from reaching its full genetically determined growth potential.
Birth weight is also related to parity. On average, the birth weight of infants born to
primiparous women is lower than of infants delivered by multiparous women (Hinkle
et al. 2014). While the biological etiology for that ﬁnding is not clear, the reason might be
the variation in the eﬃciency of materno-fetal exchange, which diﬀers between primipar-
ous and multiparous women (Hafner et al. 2013). The uteroplacenta blood ﬂow is greater
during subsequent pregnancies than it is during pregnancies involving a nulligravid uterus
(Zalud and Shaha 2008), which could result in greater observed birth weight among
infants of multiparous women.
Greater umbilical blood ﬂow means that there is not only more eﬃcient transport of
nutrients and oxygen to the fetus but also more eﬃcient transport of pollutants, especially
if the maternal environment is highly polluted. This suggests that multiparous women
may be more susceptible to the eﬀects of prenatal exposure to air pollutants than are
primiparous women. That is, greater uteroplacental blood ﬂow in multiparous women
could actually have a harmful eﬀect on fetal development, particularly in multiparous
women. The question thus arises whether and how parity conditions the risk of LBW
resulting from exposure to detrimental factors such as air pollution.
Our study sought to assess whether susceptibility to the eﬀects of prenatal exposure to
air pollutants varies between multiparous and primiparous women. It did so by testing
whether multiparous women are more prone to having infants with LBW after suﬀering
the physiological stress induced by air pollution than are primiparous women. We also
analyzed the eﬀect of air pollutants on the odds ratio for LBW separately for each parity
group. We further tested how parity conditions the observed association between exposure
to pollution and oﬀsprings’birth weight by adding an interaction term (air pollutant x
parity) into the full logistic regression model. Additionally, in order to verify if exposure
during early or late pregnancy is more important, we conducted trimester-speciﬁc
The study was conducted on a sample of 74,327 singleton, live, full-term infants
delivered between 1995 and 2009 by mothers whose residence at the time of infant’s
birth was the city of Krakow, Poland. Krakow is a densely populated metropolitan area
with unfavorable topographic locations where air pollutants are likely to accumulate.
Krakow has been among the cities most seriously threatened by air pollution not only
in Poland but also in Europe. Despite the remedial eﬀorts that have been made, Poland is
among the European countries that contributes the most (i.e., more than 10 percent) to
72 A. MERKLINGER-GRUCHALA ET AL.
the atmospheric emissions of several key pollutants: sulfur oxides, particulate matters, and
carbon monoxide, among others (European Environment Agency 2014).
Since the risk of having an infant with LBW was found to vary according to bio-
socio-economic factors such as maternal age (Reichman and Teitler 2006), parity,
marital status, education, and occupational status (Aliyu et al. 2005;Reimeetal.
2006; Shah, Zao, and Ali 2011) all of those factors were included in the analysis as
The study was carried out on a sample of live, full-term infants who were delivered between
October 1, 1995 and the end of December 2009 to mothers whose residence at the time of
infant’s birth was the city of Krakow. Birth data were obtained from the birth registry
(Central Statistical Oﬃce in Poland) and included: month and year of birth, birth weight (in
grams), sex of the oﬀspring, maternal age (in years), gestational age (in weeks), parity,
maternal education (primary, lower secondary, basic vocational, upper general or specialized
secondary, and college education), maternal employment status (employed versus not
employed), and maternal marital status (married versus not married, that is single,
widow, divorced, or separated). Due to the fact that the date of birth in registration records
is restricted to month and year, the day of birth had to be assigned as 15th of a given month
for each newborn. Season of birth was deﬁned as “heating”(for neonates born between
October and March) and “non-heating”(for those delivered between April and September).
Low birth weight (LBW) was deﬁned as birth weight below 2,500 g, and all infants with
birth weight ≥2,500 g were classiﬁed as having normal birth weight (NBW).
We conducted the research on full-term live births (gestational age 37–41 weeks) in order
to assess relationships in a more homogeneous sample and also to test whether ambient
pollutant levels could retard intrauterine growth. Such restriction enables us to separate eﬀects
of retarded fetal growth from other pathophysiological causes of low birth weight that
characterize small but preterm infants (Ma and Finch 2010).
To assess exposure to ambient air pollution we used municipal ecological monitoring
data. Daily data for carbon monoxide (CO [mg/m
]) and sulfur dioxide (SO
January 1995 to December 2009 were available from the ambient air quality monitoring
network, the system maintained by the Voivodship Inspectorate for Environmental
Protection in Krakow. In 1998 its laboratory was accredited by the Polish Centre for
Accreditation for air quality monitoring testing as the ﬁrst air quality monitoring network
laboratory in Poland. The Krakow air monitoring network provides automatic continuous
measurement of air pollutants. The approximate locations of the six monitoring stations
that provided pollution data during the study period are presented in Figure 1.
We calculated the daily averages of CO and SO
on the basis of the data from the six
monitoring stations located in the city of Krakow. Among 5,479 daily averages collected
from 1 January 1995 to 31 December 2009, only 126 (2.3 percent) daily averages of CO and
160 (2.9 percent) of SO
were missing. By using recorded gestational age and date of birth,
we estimated the weekly averages of CO and SO
levels in each pregnancy. Than we
calculated the completeness of weekly averages of CO and SO
concentrations for each
pregnancy (in percentages) by dividing the number of weekly averages of each air pollutant
during a given pregnancy by the gestational age (in weeks) and multiplying by 100. The
BIODEMOGRAPHY AND SOCIAL BIOLOGY 73
mean levels of each pollutant during an entire pregnancy (pregnancy averages) were
calculated, but only when all weekly averages of given pollutant were available for the entire
pregnancy period (that is, when the completeness of the weekly averages of given pollutant
equalled 100 percent). All other pregnancies were removed from further analyses. Thus, the
CO-pregnancy averages and SO
-pregnancy averages were calculated for 90.9 percent and
86.1 percent of neonates respectively. Such restriction was done in order to maximize the
number of weekly averages of pollutants needed to compute the mean exposure to air
pollutants during pregnancy and thus reduce bias associated with seasonal variation.
The birth registry data included 88,476 singleton newborns, from which we
excluded stillbirths (n= 387), those with unknown gestational age (n=2),those
born below 37 or above 41 weeks of gestational age (n= 7,529), those with extremely
high or low birth weight for a given gestational age (n= 5), and those born to mothers
with unknown parity (n= 5). These exclusions left 80,548 eligible subjects. The
number of cases were further restricted to those for whom the weekly averages of
either CO or SO
was 100 percent complete, which left 74,327 subject to analyses, out
of which there were n= 15,08 (2.03 percent) neonates with LBW.
Women were divided into two categories according to their employment status:
employed and not employed. We also divided women into two categories according to
the level of education: higher education (passed at least ﬁnalhighschoolexams,which
corresponds to British A-levels, such as upper general or specialized secondary and
college education) and lower education (secondary education without ﬁnal high school
Figure 1. The approximate location of the six monitoring stations that provided air pollution data
during the study period: CENTRAL station located in the Main Square of Krakow, TRAF station situated
on a busy road, representing the traﬃc zone, URB station, representing the northern urban background
site, IND station located in Nowa Huta eastern district, characterizing both industrial and suburban
zones, and SUB1 and SUB2 stations, representing southern urban background site.
74 A. MERKLINGER-GRUCHALA ET AL.
exams, such as primary education, lower secondary, and basic vocational education).
Because maternal employment status and education were strongly associated, we calcu-
lated what we called the Emp & Edu indicator, which allowed categorizing the women
into four groups: “Not Employed—Low Education”(NE-LE), “Not Employed—High
Education”(NE-HE), “Employed—Low Education”(E-LE), and “Employed—High
Education”(E-HE), with the last one being the reference group. We also stratiﬁed
women into two categories according to the parity: primiparous, i.e., women with no
previous births (reference category) and multiparous, i.e., women with one or more
The diﬀerence in mean levels of air pollutants between the heating and non-heating
seasons was assessed by a ttest for dependent samples.
We examined the distribution of several known risk factors for LBW across diﬀerent
exposure categories in order to identify potential confounders. In addition, we tested the
distribution of characteristics of singleton term births among primi- and multiparous
mothers using one-way analysis of variation (ANOVA) for continuous variable (maternal
age) and a chi-square test for categorical variables (such as sex of the oﬀspring, marital,
employment, and educational status, Emp & Edu indicator, and season of birth).
After a descriptive analysis of the maternal-infant characteristics and their distribution
among low birth weight (LBW) and non-low birth weight (NBW) groups and among
strata of parity, we performed simple and multiple logistic regression analyses to estimate
the association of each air pollutant (CO and SO
) separately as a continuous variable with
the odds ratio (OR) for LBW.
First, only the air pollutant (CO or SO
) was entered as a predictor of LBW (crude eﬀect).
In the second stage, we entered season of birth and maternal bio-socio-economic factors such
as maternal age (continuous), sex of the infant, parity (dichotomous), marital status (dichot-
omous), and Emp & Edu groups and tested whether there was a signiﬁcant change in the log
likelihood of delivering an infant with LBW after adding these factors to the simple model.
Because the main research question of our study was to assess if multiparous women
are more susceptible to a prenatal exposure to air pollutants than primiparous women, we
conducted the same procedure after stratiﬁcation for parity. The inﬂuence of parity on the
observed associations, i.e., eﬀect modiﬁcation by parity, was further tested by adding a
product term (air pollutant x parity) into the full logistic regression model. The interaction
(eﬀect modiﬁcation) referred to the state when the eﬀect of one exposure (air pollutant)
on an outcome (LBW) diﬀers across strata of another exposure (primi- and multiparous
mothers). The interaction eﬀect was interpreted on the multiplicative scale, and it indi-
cates to what extent the eﬀect of each air pollutant diﬀers between primi- and multiparous
women (Buis 2010). The stratum-speciﬁc estimates together with a statement of statistical
signiﬁcance based on an appropriate test for interaction were provided, which is the most
frequently used approach when reporting interaction in case-control and cohort studies
(Knol et al. 2009).
Additionally, in order to explore critical windows of exposure, the same analyses were
performed for the eﬀects of average air pollution levels during the ﬁrst, second, and third
trimesters of pregnancy.
Multiple comparisons were controlled by false discovery rate (FDR), which is an
expected proportion of false positives among the tests declared signiﬁcant (false plus
true). We set the maximum acceptable FDR at the level of 0.05, and calculated the
BIODEMOGRAPHY AND SOCIAL BIOLOGY 75
FDR-adjusted pvalues at this level to assess whether the resulting pvalues were still
statistically signiﬁcant after multiple comparisons. In our study, when analyzing the
exposure during the entire pregnancy, the same eﬀect of each air pollutant was tested
three times: on the entire group of mothers and in the two subsamples of parity. This
is the reason that the adjustment was performed for three comparisons among the
results of the simple logistic regression analyses (crude eﬀects)andforfourcompar-
isons among the results of the multiple logistic regression analyses (adjusted eﬀects),
because of one additional eﬀect of interaction (air pollutant x parity). After conducting
the trimester-speciﬁc estimates, the same eﬀect of each pollutant was tested nine times
in crude models (three trimesters on the entire group of mothers, three trimesters on
primiparous mothers, and three trimesters on multiparous mothers) and ten times in
the adjusted models, because of one additional eﬀect of interaction.
The results of logistic regression analyses and the test for interaction were considered
statistically signiﬁcant when the FDR-adjusted pvalue was < 0.05. The two-stage shar-
pened method was used to calculate the FDR-adjusted pvalues (Benjamini, Krieger, and
Yekutieli 2006). We used Statistica version 12.0 for the statistical analyses.
Pollutants in Krakow
From 1 January 1995 to 31 December 2009 daily concentrations of CO and SO
from 0.2 to 8.3 mg/m
, and from 1.0 to 218.5 µg/m
, respectively, with a daily average of
(SD = 0.80) and 20.0 µg/m
(SD = 20.38), respectively. The average levels of CO
during pregnancy were strongly correlated (Pearson’s correlation coeﬃcient,
r= 0.94; p< 001). The variation in both air pollutant concentrations throughout a year
was noted as similar to that found in many previous studies. In heating season the mean
levels of CO and SO
were higher than those in non-heating months (by 0.6 mg/m
, respectively; p< 0.001, Figure 2).
The univariate analysis revealed that the OR for LBW was signiﬁcantly associated with
parity, sex of the oﬀspring, maternal educational and employment status, and marital status,
but was not with maternal age (Table 1). Not married, not employed mothers and those
with lower educational status had a 94 percent, 59 percent, and 106 percent (respectively)
increased OR for delivering an infant with LBW (Table 1). When analyzing the educational
and employment status together, women who simultaneously had lower education and no
employment had almost three times higher risk of delivering a child with LBW than
employed women with higher education. Mothers who had a male oﬀspring had 38 percent
lower OR for LBW than those who gave birth to females. Multiparous women in compar-
ison to primiparous mothers had reduced odds ratio for LBW by 12 percent.
The strata of parity diﬀered in maternal age, marital, employment and educational
status, and Emp & Edu indicator but not sex of the oﬀspring and season of birth (Table 2).
76 A. MERKLINGER-GRUCHALA ET AL.
Exposure during the entire pregnancy
Crude estimates of prenatal exposure to SO
indicated slightly increased OR for LBW
among the entire group of mothers (OR = 1.01; 95% CI 1.00–1.01), and also in the strata
of multiparous only (OR = 1.01; 95% CI 1.01–1.02; Table 3). After standardization to
season of birth and maternal bio-socio-economical factors only the eﬀect of SO
the entire group of mothers (OR = 1.01; 95% CI 1.00–1.01) and among the strata of
multiparous mothers remained statistically signiﬁcant (OR = 1.01; 95% CI 1.00–1.02).
There was no association between SO
and LBW among primiparous women. A similar
pattern was seen for exposure to CO, although the association was stronger. In crude
analysis the increased OR were seen among the entire group of mothers and multiparous
mothers alone (OR = 1.20; 95% CI 1.06–1.35; OR = 1.41; 95% CI 1.18–1.67, respectively).
Also, adjusted estimates revealed that only the eﬀect of CO among the group of all
mothers (OR = 1.16; 95% CI 1.02–1.31) and among the strata of multiparous mothers
remained statistically signiﬁcant (OR = 1.28; 95% CI 1.06–1.54). Among primiparous
mothers the associations between CO and LBW were not statistically signiﬁcant both
before and after standardization (Table 3).
The test for interaction between exposure to CO and parity on the odds ratio for LBW
was statistically signiﬁcant (FDR-adjusted p= 0.03). The eﬀect of one unit increase in CO
was almost 30 percent higher in multiparous than in primiparous women (OR for
interaction = 1.28; 95% CI 1.01–1.63). The log odds for LBW among multiparous mothers
constantly rose along with increasing CO concentrations, however only after the levels of
CO exceeded 2.4 mg/m
did the log odds became higher for multiparous than for
primiparous mothers (Figure 3). In contrast, the interaction between parity and SO
was statistically nonsigniﬁcant (FDR-adjusted p= 0.08).
Exposure in trimesters
The results of multiple logistic regressions used to assess the relationship between air
pollutant exposure in trimesters and odds ratio for LBW are presented in Table 4. Crude
estimates of prenatal exposure to SO
during the ﬁrst trimester indicated statistically
signiﬁcant but weakly increased OR for LBW among the entire group of mothers and
Figure 2. Mean monthly sulfur dioxide (SO
) and carbon monoxide (CO) levels in Krakow city across the
study period (1995–2009). Winter levels are elevated in comparison with summer levels due to the
heating season (p< 0.001).
BIODEMOGRAPHY AND SOCIAL BIOLOGY 77
also in the strata of multiparous only mothers. After standardization, the pattern of
observed associations remained the same, that is, among all mothers (OR = 1.01; 95%
CI 1.00–1.01) and among multiparous mothers (OR = 1.01; 95% CI 1.00–1.01) we found
weakly elevated odds ratio for LBW. Exposure to SO
during the second and third
trimesters was not associated with LBW either in the group of all mothers or in either
strata of parity equally before and after standardization.
Likewise, exposure to CO during the ﬁrst trimester both before and after adjustment
was associated with OR for LBW, and the relationships were stronger than for SO
Adjusted estimates for CO exposure during this trimester were associated with 16%
increased OR for LBW among the group of all mothers (95% CI 1.03–1.30) and with
Table 1. Distribution of characteristics of singleton, full-term births by low birth weight status.
Characteristic Level NBW LBW OR (95% CI) pvalue
Sex of the oﬀspring Girl n35,331 908 1.00 [reference]
% 48.5 60.2
Boy n37,488 600 0.62 (0.56–0.69) < 0.001
% 51.5 39.8
Parity Primiparous n39,355 864 1.00 [reference]
% 54.0 57.3
Multiparous n33,464 644 0.88 (0.79–0.97) < 0.001
% 46.0 42.7
Marital status Married n62,854 1153 1.00 [reference]
% 86.3 76.5
Not married n9,965 355 1.94 (1.72–2.19) < 0.001
% 13.7 23.5
Employment status Employed n 54,321 974 1.00 [reference]
% 74.8 65.2
Not employed n18274 520 1.59 (1.43–1.77) < 0.001
% 25.2 34.8
Educational status Higher n56,759 947 1.00 [reference]
% 78.2 63.5
Lower n15,830 545 2.06 (1.85–2.30) < 0.001
% 21.8 36.5
Emp & Edu* E-HE n45,836 753 1.00 [reference]
% 63.2 50.5
NE-HE n10,895 194 1.09 (0.92–1.27) 0.321
% 15.0 13.0
E-LE n8,451 219 1.57 (1.35–1.84) < 0.001
% 11.6 14.7
NE-LE n7,364 325 2.69 (2.35–3.07) < 0.001
% 10.2 21.8
Maternal age (year) n72,819 1508 1.00 (0.99–1.01) 0.968
Mean 28.1 28.1
SD 5.2 5.9
Season of birth Non-heating n37,018 729 1.00 [reference]
% 50.8 48.3
Heating n35,801 779 1.10 (1.00–1.22) 0.055
% 49.2 51.7
)n71,750 1,480 1.20 (1.06–1.35) 0.003
Mean 1.3 1.4
SD 0.4 0.4
)n67,942 1,410 1.01 (1.00–1.01) 0.009
Mean 20.4 21.1
SD 10 10
Note. *Emp & Edu indicator: NE-LE = Not Employed—Low Education, NE-HE = Not Employed—High Education, E-LE =
Employed—Low Education, E-HE = Employed—High Education.
78 A. MERKLINGER-GRUCHALA ET AL.
25% increased OR for LBW among multiparous only mothers (95% CI 1.05–1.49),
however these results were of only borderline signiﬁcance (Table 4).
A statistically signiﬁcant relationship between exposure to CO during the third trime-
ster of pregnancy and increased OR for LBW among the group of all mothers and among
the strata of multiparous mothers observed before adjustment did not reach statistical
signiﬁcance after standardization.
Because adjusted models of analyses in trimesters revealed that both pollutants increased
OR for LBW among the entire group of mothers and among the strata of multiparous
mothers but only when exposure occurs during the ﬁrst trimester of pregnancy, we further
tested this subgroup eﬀects by introducing the interaction term (air pollutant in ﬁrst
trimester x parity) into the full models. Although increased exposure to SO
ﬁrst trimester was associated (after standardization) with LBW among multiparous mothers
and not in primiparous mothers, the test for interaction revealed that these subgroup
diﬀerences referring to this particular time of pregnancy are not statistically signiﬁcant
(FDR-adjusted pfor interaction = 0.15). Similarly, the interaction between exposure to CO
in the ﬁrst trimester and parity was nonsigniﬁcant (FDR-adjusted pfor interaction = 0.57).
Table 2. Distribution of characteristics of singleton, full-term births stratiﬁed for parity.
Characteristic Level Multiparous Primiparous pvalue*
Sex of the oﬀspring (n, %) Girl 16,752 49.1% 19,487 48.5% 0.072
Boy 17,356 50.9% 20,732 51.5%
Maternal age (mean, SD) 30.5 4.8 26.1 4.6 < 0.001
Marital status (n, %) Married 30,715 90.1% 33,292 82.8% < 0.001
Not married 3,393 9.9% 6,927 17.2%
Employment status (n, %) Employed 25,088 73.9% 30,207 75.3% < 0.001
Not employed 8,865 26.1% 9,929 24.7%
Educational status (n, %) Higher 25,201 74.2% 32,505 81.0% < 0.001
Lower 8,758 25.8% 7,617 19.0%
Emp & Edu** (n, %) E-HE 20,363 60.0% 26,226 65.4% < 0.001
NE-HE 4,826 14.2% 6,263 15.6%
E-LE 4,713 13.9% 3,957 9.9%
NE-LE 4,034 11.9% 3,655 9.1%
Season of birth (n, %) Heating 16,685 48.9% 19,895 49.5% 0.136
Non-heating 17,423 51.1% 20,324 50.5%
Note. *One-way ANOVA was used for continuous and chi-square test for categorical variables.
**Emp & Edu indicator: NE-LE = Not Employed—Low Education, NE-HE = Not Employed—High Education,
E-LE = Employed—Low Education, E-HE = Employed—High Education.
Table 3. Crude and adjusted odds ratios (OR) with 95% conﬁdence intervals (CI) for low birth weight
(LBW) by pollution exposure and parity.
Adjusted to bio-socio-economical factors and season of
OR FDR-adjusted pvalue OR FDR-adjusted pvalue
All 1.20 (1.06–1.35) < 0.01 1.16 (1.02–1.31) 0.02
Primiparous 1.06 (0.90–1.24) 0.18 1.02 (0.86–1.21) 0.41
Multiparous 1.41 (1.18–1.67) < 0.01 1.28 (1.06–1.54) < 0.01
All 1.01 (1.00–1.01) 0.01 1.01 (1.00–1.01) 0.02
Primiparous 1.00 (1.00–1.01) 0.18 1.00 (0.99–1.01) 0.38
Multiparous 1.01 (1.01–1.02) < 0.01 1.01 (1.00–1.02) < 0.01
BIODEMOGRAPHY AND SOCIAL BIOLOGY 79
0,4 0,6 0,8 1,0 1,2 1,4 1,6 1,8 2,0 2,2 2,4 2,6 2,8
Log odds for LBW
Figure 3. Log odds for LBW, as a function of exposure to CO according to parity, after controlling for
potential confounding factors (i.e., season of birth, sex of the oﬀspring, marital status, employment
status, education, and maternal age), with assumption that all predictors are unchanged, that is, set at
the reference level (categorical variables), while maternal age (continuous) is held ﬁxed at 30 years.
Table 4. Crude and adjusted odds ratios (OR) with 95% conﬁdence intervals (CI) for low birth weight
(LBW) by pollution exposure, parity, and trimester.
Adjusted to bio-socio-economical factors and
season of birth
OR FDR-adjusted pvalue OR FDR-adjusted pvalue
Trimester 1 1.14 (1.04–1.25) 0.01 1.16 (1.03–1.30) 0.05
Trimester 2 1.06 (0.97–1.17) 0.17 0.99 (0.83–1.18) 0.93
Trimester 3 1.14 (1.04–1.25) 0.01 1.03 (0.87–1.21) 0.93
Trimester 1 1.07 (0.95–1.22) 0.21 1.09 (0.93–1.27) 0.57
Trimester 2 1.00 (0.88–1.14) 0.55 1.02 (0.80–1.29) 0.93
Trimester 3 1.01 (0.89–1.15) 0.55 0.93 (0.74–1.17) 0.91
Trimester 1 1.23 (1.07–1.41) 0.01 1.25 (1.05–1.49) 0.05
Trimester 2 1.15 (1.00–1.32) 0.05 0.95 (0.73–1.24) 0.93
Trimester 3 1.31 (1.15–1.50) < 0.01 1.16 (0.90–1.48) 0.57
Trimester 1 1.00 (1.00–1.01) 0.03 1.01 (1.00–1.01) 0.01
Trimester 2 1.00 (1.00–1.00) 0.69 1.00 (0.99–1.00) 0.45
Trimester 3 1.00 (1.00–1.01) 0.06 1.00 (1.00–1.01) 0.15
Trimester 1 1.00 (1.00–1.01) 0.60 1.00 (1.00–1.01) 0.45
Trimester 2 1.00 (0.99–1.00) 0.71 1.00 (0.99–1.00) 0.45
Trimester 3 1.00 (1.00–1.01) 0.56 1.00 (1.00–1.01) 0.45
Trimester 1 1.01 (1.00–1.01) < 0.01 1.01 (1.00–1.02) < 0.01
Trimester 2 1.00 (1.00–1.01) 0.58 1.00 (0.99–1.01) 0.61
Trimester 3 1.01 (1.00–1.01) 0.06 1.01 (1.00–1.01) 0.15
80 A. MERKLINGER-GRUCHALA ET AL.
We have found that 1 mg/m
increase in carbon monoxide exposure during the entire
pregnancy was associated with a 16 percent higher odds ratio for low birth weight after
standardization to season of birth and maternal bio-socio-economic factors. Similarly, the
adjusted odds ratio for low birth weight was 1 percent higher after increasing exposure to
sulfur dioxide by 1 µg/m
during the entire pregnancy period.
These results are in accordance with previous ﬁndings showing lower birth weight or
increased risk of LBW in term neonates whose mothers were exposed to elevated levels of
ambient CO (Lin et al. 2004; Morello-Frosch et al. 2010)orSO
during the entire
pregnancy (Bobak and Leon 1999; Geer, Weedon, and Bell 2012; Lin et al. 2004).
Considering speciﬁc windows of exposure, our study conﬁrmed previous ﬁndings of
increased risk after exposure to SO
during the ﬁrst trimester of pregnancy (Bobak
2000; Ha et al. 2001) but not those where elevated risk was linked with the last trimester
(Lin et al. 2004). Correspondingly, our results are in accordance with those studies that
showed an association between low birth weight and exposure to CO during the ﬁrst
trimester (Ha et al. 2001) but not during the last trimester of pregnancy (Ritz and Yu
1999). Furthermore, we noticed that the eﬀect of exposure to CO during the entire
pregnancy on the birth weight of neonates is not the same in all mothers, but diﬀers
according to their parity, which was conﬁrmed by appropriate tests for interactions. The
harmful eﬀect of CO was shown among the multiparous mothers: increase by 1 mg/m
CO was associated with 28 percent elevated odds ratio for LBW, while increasing exposure
to CO among primiparous mothers was not associated with higher risk. Our results
suggest that multiparous mothers may be more vulnerable to carbon monoxide than
primiparous mothers, which is reﬂected in the higher probability of having a child with
birth weight below 2,500 g at term. Thus, we propose that parity may be a modiﬁer of the
association between CO and the risk of LBW, but further studies are warranted.
The ﬁnding that the oﬀspring of multiparous women are more strongly aﬀected by
ambient CO levels during pregnancy than the infants born to primiparous mothers
has been reported in only one previous study conducted in Los Angeles, California
(Ritz and Yu 1999). The eﬀect of CO on birth weight was reported only for higher
parity mothers, that is, those giving birth to a second or higher-order child. High-
parity women in the ≥95th percentile of exposure to CO concentrations in compar-
isontowomeninthe≤50th percentile for last trimester exposure had almost two-
times higher odds ratio for LBW, whereas similar exposure among the entire group
of women was not associated with LBW.
The eﬀect-measure of other chemicals having an impact on birth outcomes can also be
modiﬁed by parity. For example, perﬂuorooctanesulfonate (PFOS) and perﬂuorooctanoate
(PFOA), synthetic compounds or metabolites of other perﬂuorinated chemicals that are
used in a variety of consumer products, such as nonstick pans, carpets, furniture, and
household cleaners, were associated with the ponderal index of neonates, but in the
opposite direction in the two strata of parity (Fei et al. 2008). Maternal exposure to
PFOS and PFOA were associated negatively with the ponderal index among multiparous
women but positively among nulliparous women.
It is biologically plausible that air pollutants and other toxins negatively aﬀect birth
outcomes, especially among multiparous mothers. Parity may manipulate the
BIODEMOGRAPHY AND SOCIAL BIOLOGY 81
development of the placenta and placental eﬃciency (Roland et al. 2012; Wilsher and
Allen 2003), which can be linked not only with nutrient but also toxin exchange. Greater
uteroplacental blood ﬂow among multiparous than primiparous mothers (Zalud and
Shaha 2008) and better placental bed perfusion due to improvement of the placental
bed vascularization (Hafner et al. 2013) have been shown. In addition, a ﬁrst pregnancy
was found to result in pioneering structural changes of the internal elastic lamina in spiral
arteries, and these modiﬁcations were not totally eliminated following parturition (Khong,
Adema, and Erwich 2003). The authors suggested that these residual arterial alterations
allow trophoblasts to remodel arteries more eﬀectively in a subsequent pregnancy.
Moreover, total antioxidant capacity has been shown to be lower and oxidative stress
indicators higher in cord blood during second or subsequent pregnancies than in the null
gravid uterus (Mutlu et al. 2012). All of these morphological and physiological amend-
ments following a ﬁrst pregnancy may be responsible for the higher susceptibility of the
fetuses of multiparous mothers to environmental pollutants. It seems reasonable to
assume that when the level of CO is low, the more eﬃcient placental blood circulation
in multiparous mothers protects them from having LBW oﬀspring by assuring enough
nutrient and oxygen supply to the fetus. But during elevated pollutant concentrations
higher umbilical blood ﬂow may be deleterious and outweighs the advantages of this
adaptive strategy seen when the pollution is low.
The relationship between high parity and placental eﬃciency could also potentially
result indirectly from low socioeconomic status (SES) of the mother. Being multiparous
may be connected with poor SES and lower education (Roman et al. 2004), which in turn
can be linked with dietary calorie and/or protein deprivation (Bojar et al. 2007). Maternal
malnutrition inﬂuences both the maternal and fetal somatotrophic axis, while nutritional
and endocrine signals can increase placental transport capacity (Fowden et al. 2009).
Many studies revealed that in mothers from nutritionally deprived populations, but also
in well-nourished women from developed countries, each subsequent pregnancy deterio-
rates the maternal condition, which may result in poor pregnancy outcome. This phe-
nomenon associated with high lifetime costs of reproduction is called the “maternal
depletion syndrome”(Dewey and Cohen 2007; King 2003; van Eijsden et al. 2008; Zhu
et al. 1999) and may have long-term health consequences for the mother, such as serious
metabolic disorders or even a reduced life span (Jasienska 2013).
Our study, just like all studies based on data registries, was not able to take into considera-
tion individual characteristics of women, such as dietary or smoking habits, that might
provide additional information important to consider when assessing relationships between
air pollution and LBW risk. Moreover, air pollution data based on community monitoring
systems do not provide information about variability among individuals’exposure due to
occupational exposure or outdoor/indoor activity patterns. Moreover, because of no precise
geographical information on the residency of the women, we can look only at temporal
variation in air pollution and not take into consideration within-city variability in pollution
concentrations. But when considering spatial resolution assessment based on regulatory
monitoring network data we noted that the number of monitoring stations per land area in
Krakow is high (almost 6/300 km
) in comparison to other study locations, for example,
California’s South Coast Air Basin (0.6/300 km
) and even Vancouver (1.5/300 km
was regarded to have very high monitor density by Marshall, Nethery, and Brauer (2008). The
high quality of the network of air pollution stations in Malopolska province and its
82 A. MERKLINGER-GRUCHALA ET AL.
appropriate coverage of the Krakow residency area (six stations per 32685 ha in a city of
almost 760,000 inhabitants; Central Statistical Oﬃce 2013) was also conﬁrmed by a cohort
study conducted on pregnant women, the residents of the city of Krakow (Jedrychowski et al.
Whilesomeauthorsassignexposurelevels according to the monitoring station
closest to each mother’s residence, the information about residence of the women
who participated in the study was not available to us and therefore we were not able to
apply modeling techniques (such as LUR models) that join together air pollution and
geographic information to characterize the variability in exposure. However, pregnant
womenlivinginmetropolitanlocationsare not exposed to pollutants only in their
place of residence, but also in their workplace, during their travel to work, and during
all other activities (shopping, visits to relatives and friends, recreational activities, etc.),
and thus their exposure might be better represented by the average, rather than by just
one station. This problem was also discussed by others (Gouveia, Bremner, and Novaes
2004). We are aware that assessing exposure in this way is connected with potential
error, but, especially given our large sample size, it is likely to be nondiﬀerential, i.e.,
to have the same magnitude and direction among LBW and NBW neonates. Moreover,
the most serious consequence of such an error is the attenuation of the eﬀect estimated
(Kreienbrock 2007). Further, when considering the estimation of exposure for popula-
tions close to the nearest monitor’s location in epidemiological settings, a distance of
50 km is very often chosen as a reasonable distance for extrapolation of observed air
pollutant concentrations (Bravo et al. 2012;O’Donnell et al. 2011; Spencer-Hwang
et al. 2011). The Krakow city area has a size of 18 km x 31 km and within this area
there are six monitoring stations, thus it seems rational to assume that the estimation
of exposure in our study population is reliable.
Another approach to assess individual exposure by asking participants to carry a
pollution-measuring device is a much better tool, however, it does not allow analyzing a
large sample of mothers and is unlikely that such a selected study group would be
representative of the entire population.
In summary, our research conducted on a very large population of infants suggests that
the odds ratio for LBW may vary with respect to parity and that multiparous mothers are
more susceptible to unfavorable prenatal conditions such as exposure to carbon mon-
oxide. These ﬁndings provide additional evidence for the relationship between air pollu-
tion and LBW by identifying a vulnerable subgroup of mothers.
The ubiquity and persistence of air pollutants pose unique and emerging challenges in
addressing this public health issue. State and local public health professionals together
with reproductive health providers need to focus their eﬀorts on reaching subpopulations
who are at high risk for prenatal exposure to environmental pollutants. One such
disadvantaged subgroup may be multiparous pregnant women.
Disclosure of potential conﬂicts of interest
The authors declare that they have no conﬂict of interest. This article does not contain any studies
with human participants performed by any of the authors.
BIODEMOGRAPHY AND SOCIAL BIOLOGY 83
This study was carried out within the Framework of a Project No. N N404 055 136, ﬁnanced by the
Ministry of Science and Higher Education.
Aliyu, M. H., P. E. Jolly, J. E. Ehiri, and H. M. Salihu. 2005. High parity and adverse birth outcomes:
Exploring the maze. Birth 32 (1):45–59. doi:10.1111/j.0730-7659.2005.00344.x.
André, L., F. Gouzi, J. Thireau, G. Meyer, J. Boissiere, M. Delage, A. Abdellaoui, C. Feillet-Coudray,
G. Fouret, J.-P. Cristol, A. Lacampagne, P. Obert, C. Reboul, J. Fauconnier, M. Hayot, S. Richard,
and O. Cazorla. 2011. Carbon monoxide exposure enhances arrhythmia after cardiac stress:
Involvement of oxidative stress. Basic Research in Cardiology 106 (6):1235–46. doi:10.1007/
Aubard, Y., and I. Magne. 2000. Carbon monoxide poisoning in pregnancy. BJOG: An International
Journal of Obstetrics & Gynaecology 107 (7):833–38. doi:10.1111/bjo.2000.107.issue-7.
Benjamini, Y., A. Krieger, and D. Yekutieli. 2006. Adaptive linear step-up procedures that control
the false discovery rate. Biometrika 93:491–507. doi:10.1093/biomet/93.3.491.
Bobak, M. 2000. Outdoor air pollution, low birth weight, and prematurity. Environmental Health
Perspectives 108 (2):173–76. doi:10.1289/ehp.00108173.
Bobak, M., and D. A. Leon. 1999. Pregnancy outcomes and outdoor air pollution: An ecological
study in districts of the Czech Republic 1986-8. Occupational and Environmental Medicine 56
Bojar, I., E. Humeniuk, L. Wdowiak, P. Miotła, E. Warchoł-Sławińska, and K. Włoch. 2007.
Nutritional behaviours of pregnant women. Problemy Higieny I Epidemiologii 88:74–77.
Bravo, M. A., M. Fuentes, Y. Zhang, M. J. Burr, and M. L. Bell. 2012. Comparison of exposure
estimation methods for air pollutants: Ambient monitoring data and regional air quality simula-
tion. Environmental Research 116:1–10. doi:10.1016/j.envres.2012.04.008.
Buis, M. L. 2010. Stata tip 87: Interpretation of interactions in nonlinear models. Stata Journal
Central Statistical Oﬃce. 2013. Statistical Yearbook of the Republic of Poland, 2013 (accessed
October 19, 2015 http://stat.gov.pl/cps/rde/xbcr/gus/RS_rocznik_statystyczny_rp_2013.pdf.
Dewey, K. G., and R. J. Cohen. 2007. Does birth spacing aﬀect maternal or child nutritional status?
A systematic literature review. Maternal & Child Nutrition 3(3): 151–73. Review. doi: 10.1111/
European Environment Agency. 2014. Air quality in Europe —2014 report. Luxembourg:
Publications Oﬃce of the European Union. http://www.eea.europa.eu/publications/air-quality-
in-europe-2014/download (accessed October 19, 2015).
Fei, C., J. K. McLaughlin, R. E. Tarone, and J. Olsen. 2008. Fetal growth indicators and perﬂuori-
nated chemicals: A study in the Danish National Birth Cohort. American Journal of Epidemiology
168 (1):66–72. doi:10.1093/aje/kwn095.
Fowden, A. L., A. N. Sferruzzi-Perri, P. M. Coan, M. Constancia, and G. J. Burton. 2009. Placental
eﬃciency and adaptation: Endocrine regulation. Journal of Physiology 587 (14):3459–72.
Geer, L. A., J. Weedon, and M. L. Bell. 2012. Ambient air pollution and term birth weight in Texas
from 1998 to 2004. Journal of the Air & Waste Management Association 62 (11):1285–95.
Gouveia, N., S. A. Bremner, and H. M. D. Novaes. 2004. Association between ambient air pollution
and birth weight in São Paulo, Brazil. Journal of Epidemiology and Community Health 58 (1):11–
Ha, E. H., Y. C. Hong, B. E. Lee, B. H. Woo, J. Schwartz, and D. C. Christiani. 2001. Is air pollution
a risk factor for low birth weight in Seoul? Epidemiology (Cambridge, Mass.) 12 (6):643–8.
84 A. MERKLINGER-GRUCHALA ET AL.
Hafner, E., M. Metzenbauer, I. Stumpﬂen, and T. Waldhor. 2013. Measurement of placental bed
vascularization in the ﬁrst trimester, using 3D-power-Doppler, for the detection of pregnancies
at-risk for fetal and maternal complications. Placenta 34 (10):892–98. doi:10.1016/j.
Hinkle, S. N., P. S. Albert, P. Mendola, L. A. Sjaarda, E. Yeung, N. S. Boghossian, and S. K. Laughon.
2014. The association between parity and birthweight in a longitudinal consecutive pregnancy
cohort. Paediatric and Perinatal Epidemiology 28 (2):106–15. doi:10.1111/ppe.12099.
Jasienska, G. 2013. The fragile wisdom: An evolutionary view on women’s biology and health.
Cambridge, MA: Harvard University Press.
Jasienska, G., A. Ziomkiewicz, I. Thune, S. F. Lipson, and P. T. Ellison. 2006. Habitual physical
activity and estradiol levels in women of reproductive age. European Journal of Cancer Prevention
15 (5):439–45. doi:10.1097/00008469-200610000-00009.
Jedrychowski, W., I. Bendkowska, E. Flak, A. Penar, R. Jacek, I. Kaim, J. D. Spengler, D. Camann,
and F. P. Perera. 2004. Estimated risk for altered fetal growth resulting from exposure to ﬁne
particles during pregnancy: An epidemiologic prospective cohort study in Poland. Environmental
Health Perspectives 112 (14):1398–402. doi:10.1289/ehp.7065.
Johnson, F. K., and R. A. Johnson. 2003. Carbon monoxide promotes endothelium-dependent
constriction of isolated gracilis muscle arterioles. American Journal of Physiology. Regulatory,
Integrative and Comparative Physiology 285 (3):R536–41. doi:10.1152/ajpregu.00624.2002.
Kapiszewska, M., M. Miskiewicz, P. T. Ellison, I. Thune, and G. Jasienska. 2006. High tea con-
sumption diminishes salivary 17β-estradiol concentration in Polish women. British Journal of
Nutrition 95 (5):989–95. doi:10.1079/BJN20061755.
Khong, T. Y., E. D. Adema, and J. J. H. M. Erwich. 2003. On an anatomical basis for the increase in
birth weight in second and subsequent born children. Placenta 24 (4):348–53. doi:10.1053/
King, J. C. 2003. The risk of maternal nutritional depletion and poor outcomes increases in early or
closely spaced pregnancies. Journal of Nutrition 133 (5):1732S–1736S.
Knol, M. J., M. Egger, P. Scott, M. L. Geerlings, and J. P. Vandenbroucke. 2009. When one depends
on the other: Reporting of interaction in case-control and cohort studies. Epidemiology 20
Kreienbrock, L. 2007. Environmental epidemiology. In Handbook of epidemiology, eds. W. Ahrens,
and I. Pigeot, 951–998. Berlin: Springer Science & Business Media.
Lin, C. M., C. Y. Li, G. Y. Yang, and I. F. Mao. 2004. Association between maternal exposure to
elevated ambient sulfur dioxide during pregnancy and term low birth weight. Environmental
Research 96 (1):41–50. doi:10.1016/j.envres.2004.03.005.
Ma, S., and B. K. Finch. 2010. Birth outcome measures and infant mortality. Population Research
and Policy Review 29 (6):865–91. doi:10.1007/s11113-009-9172-3.
Marshall, J. D., E. Nethery, and M. Brauer. 2008. Within-urban variability in ambient air pollution:
Comparison of estimation methods. Atmospheric Environment 42 (6):1359–69. doi:10.1016/j.
Merklinger-Gruchala, A., P. T. Ellison, S. F. Lipson, I. Thune, and G. Jasienska. 2008. Low estradiol
levels in women of reproductive age having low sleep variation. European Journal of Cancer
Prevention 17 (5):467–72. doi:10.1097/CEJ.0b013e3282f75f67.
Merklinger-Gruchala, A., and M. Kapiszewska. 2015. Association between PM10 air pollution and
birth weight after full-term pregnancy in Krakow city 1995-2009—Trimester speciﬁcity. Annals of
Agricultural and Environmental Medicine 22 (2):265–70. doi:10.5604/12321966.1152078.
Mohorovic, L., and V. Micovic. 2012. The importance of “ﬁrst-blood circulation stage”, a new
insight into the pathogenesis of clinical manifestations of preeclampsia. Advances in Bioscience
and Biotechnology 3 (7A). doi:10.4236/abb.2012.327116.
Morello-Frosch, R., B. M. Jesdale, J. L. Sadd, and M. Pastor. 2010. Ambient air pollution exposure and
full-term birth weight in California. Environmental Health 9:44. doi:10.1186/1476-069X-9-44.
Mutlu, B., A. Y. Bas, N. Aksoy, and A. Taskin. 2012. The eﬀect of maternal number of births on
oxidative and antioxidative systems in cord blood. Journal of Maternal-Fetal & Neonatal
Medicine 25 (6):802–05. doi:10.3109/14767058.2011.594920.
BIODEMOGRAPHY AND SOCIAL BIOLOGY 85
O’Donnell, M. J., J. Fang, M. A. Mittleman, M. K. Kapral, and G. A. Wellenius. 2011. Fine
particulate air pollution (PM2.5) and the risk of acute ischemic stroke. Epidemiology 22
Reichman, N., and J. Teitler. 2006. Paternal age as a risk factor for low birthweight. American
Journal of Public Health 96.5:862–66. doi:10.2105/AJPH.2005.066324.
Reime, B., P. A. Ratner, S. N. Tomaselli-Reime, A. Kelly, B. A. Schuecking, and P. Wenzlaﬀ. 2006.
The role of mediating factors in the association between social deprivation and low birth weight
in Germany. Social Science & Medicine 62 (7):1731–44. doi:10.1016/j.socscimed.2005.08.017.
Ritz, B., and F. Yu. 1999. The eﬀect of ambient carbon monoxide on low birth weight among
children born in southern California between 1989 and 1993. Environmental Health Perspectives
107 (1):17–25. doi:10.1289/ehp.9910717.
Roland, M. C., C. M. Friis, N. Voldner, K. Godang, J. Bollerslev, G. Haugen, T. Henriksen, and N.
Harvey. 2012. Fetal growth versus birthweight: The role of placenta versus other determinants.
PLoS One 7 (6):e39324. doi:10.1371/journal.pone.0039324.
Roman, H., P. Y. Robillard, E. Verspyck, T. C. Hulsey, L. Marpeau, and G. Barau. 2004. Obstetric
and neonatal outcomes in grand multiparity. Obstetrics and Gynecology 103 (6):1294–99.
Rudra, C. B., M. A. Williams, L. Sheppard, J. Q. Koenig, M. A. Schiﬀ, I. O. Frederick, and R. Dills.
2010. Relation of whole blood carboxyhemoglobin concentration to ambient carbon monoxide
exposure estimated using regression. American Journal of Epidemiology 171 (8):942–51.
Shah, P. S., J. Zao, and S. Ali. 2011. Maternal marital status and birth outcomes: A systematic
review and meta-analyses. Maternal and Child Health Journal 15 (7):1097–109. doi:10.1007/
Spencer-Hwang, R., S. F. Knutsen, S. Soret, M. Ghamsary, W. L. Beeson, K. Oda, D. Shavlik, and N.
Jaipaul. 2011. Ambient air pollutants and risk of fatal coronary heart disease among kidney
transplant recipients. American Journal of Kidney Diseases 58 (4):608–16. doi:10.1053/j.
Stieb, D. M., L. Chen, M. Eshoul, and S. Judek. 2012. Ambient air pollution, birth weight and
preterm birth: A systematic review and meta-analysis. Environmental Research 117:100–11.
Tomei, G., M. Ciarrocca, B. R. Fortunato, A. Capozzella, M. V. Rosati, D. Cerratti, E. Tomao, V.
Anzelmo, C. Monti, and F. Tomei. 2006. Exposure to traﬃc pollutants and eﬀects on 17-β-
estradiol (E2) in female workers. International Archives of Occupational and Environmental
Health 80 (1):70–77. doi:10.1007/s00420-006-0105-8.
van Eijsden, M., L. J. Smits, M. F. van der Wal, and G. J. Bonsel. 2008. Association between short
interpregnancy intervals and term birth weight: The role of folate depletion. American Journal of
Clinical Nutrition 88(1): 147–53.
Wells, J. C. 2000. Natural selection and sex diﬀerences in morbidity and mortality in early life.
Journal of Theoretical Biology 202 (1):65–76. doi:10.1006/jtbi.1999.1044.
Wilsher, S., and W. R. Allen. 2003. The eﬀects of maternal age and parity on placental and fetal
development in the mare. Equine Veterinary Journal 35 (5):476–83. doi:10.2746/
Woodruﬀ, T. J., S. J. Janssen, L. J. Guillette, Jr., and L. C. Giudice, eds. 2010. Environmental impacts
on reproductive health and fertility. New York: Cambridge University Press.
Zalud, I., and S. Shaha. 2008. Three-dimensional sonography of the placental and uterine spiral
vasculature: Inﬂuence of maternal age and parity. Journal of Clinical Ultrasound 36 (7):391–96.
Zhu, B. P., R. T. Rolfs, B. E. Nangle, and J. M. Horna. 1999. Eﬀect of the interval between
pregnancies on perinatal outcome. New England Journal of Medicine 340:589–94. doi:10.1056/
Ziomkiewicz, A., P. T. Ellison, S. F. Lipson, I. Thune, and G. Jasienska. 2008. Body fat, energy
balance and estradiol levels: A study based on hormonal proﬁles from complete menstrual cycles.
Human Reproduction 23 (11):2555–63. doi:10.1093/humrep/den213.
86 A. MERKLINGER-GRUCHALA ET AL.