Background Low birth weight (LBW) is a public health problem linked to lack of equity in populations. Despite efforts to decrease the proportion of newborns with LBW, success has been quite limited. In recent years, studies focused on explaining how social factors influence this problem have shown that populations with greater inequities have a greater proportion of newborns with LBW.
Background
Low birth weight (LBW) is a public health problem linked to lack of equity in populations. Despite efforts to decrease the proportion of newborns with LBW, success has been quite limited. In recent years, studies focused on explaining how social factors influence this problem have shown that populations with greater inequities have a greater proportion of newborns with LBW.Methods The objective was to describe socioeconomic factors related to LBW adjusted by demographic, reproductive and health services variables in Mexico City. A case-control study was carried out in three hospitals with gynaecological and obstetrics services in Mexico City during the first half of 1996. During the recruiting period all children with LBW (cases), defined as newborns weighing <2500 grams, were matched with children born on the same day to control for time of birth. Upon arrival at the hospital for delivery, women were interviewed to determine if they met our inclusion criteria. Women with a history of chronic conditions and those with twin or multiple pregnancies were excluded. Variables with clinical and statistical significance were included in a multivariate model (logistic regression).
Methods
The objective was to describe socioeconomic factors related to LBW adjusted by demographic, reproductive and health services variables in Mexico City. A case-control study was carried out in three hospitals with gynaecological and obstetrics services in Mexico City during the first half of 1996. During the recruiting period all children with LBW (cases), defined as newborns weighing <2500 grams, were matched with children born on the same day to control for time of birth. Upon arrival at the hospital for delivery, women were interviewed to determine if they met our inclusion criteria. Women with a history of chronic conditions and those with twin or multiple pregnancies were excluded. Variables with clinical and statistical significance were included in a multivariate model (logistic regression).Results We found that low socioeconomic level was the most important risk factor for LBW and was independent of other factors, including those related to reproduction and nutrition, smoking, morbidity during pregnancy, accessibility to health services and prenatal care (OR 2.68; 95% CI 1.19, 6.03).
Results
We found that low socioeconomic level was the most important risk factor for LBW and was independent of other factors, including those related to reproduction and nutrition, smoking, morbidity during pregnancy, accessibility to health services and prenatal care (OR 2.68; 95% CI 1.19, 6.03).Conclusion We found that socioeconomic factors are relevant to LBW. However further research should be done in different population groups as well as developing precise ways of measuring socioeconomic factors and their role in reproductive health.
Conclusion
We found that socioeconomic factors are relevant to LBW. However further research should be done in different population groups as well as developing precise ways of measuring socioeconomic factors and their role in reproductive health.

Background On average, the worldwide incidence of low birthweight (LBW) is 17% per year, making LBW an important infant health problem in many populations [ 1 ]. The incidence of LBW varies among countries, ranging from 4% to 6% in Western countries like Sweden, France, United States and Canada (UNICEF 2003) and much higher in developing countries. In Latin America, the overall LBW rate varies according to geographical region. The Pan American Health Organization has estimated that the overall LBW rate is 8.27% in South America and Mexico and ranges from 6% in Peru to 10% in Bolivia and Venezuela. In Central America the overall LBW rate is 11.2%, ranging from 6% in Belize and El Salvador to 15% in Nicaragua [ 2 ]. The LBW rate can also vary within countries. For example, in Mexico, the incidence ranges from 8.2% to 12%, depending on geographic al region [ 3 ]. In addition to its impact on infant mortality, LBW has been associated with higher probabilities of infection, malnutrition and handicapping conditions during childhood, including cerebral palsy, mental deficiencies and problems related to behavior and learning during childhood [ 4 - 6 ]. Children who survive LBW have a higher incidence of diseases, retardation in cognitive development, and undernourishment. There is also evidence that LBW or its determinant factors are associated with a predisposition to higher rates of diabetes, cardiac diseases and other future chronic health problems [ 7 - 9 ]. The biological processes that affect the fetus  in utero  are related to the mother's physiology, including her nutrition (mother's weight before pregnancy and history of having newborns with LBW), exercise, infections, and consumption of tobacco, alcohol and other drugs [ 10 , 11 ]. During the fetal phase, growth depends on the nutritional condition of the mother, indicating that pregnant women should not only increase their weight but also consume essential nutrients. For many women in the developing world, however, economic, social and cultural factors make it difficult for them to obtain the necessary food and healthcare, which are closely interrelated [ 12 ]. Associations between poorer child health and poverty, inequity and social exclusion have been documented worldwide and have been shown to be independent of research methods, local culture, and available health care services [ 13 ]. While the relationship between socio-economic conditions and health have been of interest and concern for centuries, recent studies have sought to identify the social factors most relevant to health [ 14 ]. Some authors consider that health therefore may be an important determinant of opportunities in life and this process, termed "selection by health", and suggests that health "selects" people in different social strata [ 15 , 16 ]. Among the socio-economic factors are income, education, occupation, household leadership and gender differences related to roles within the family [ 17 , 18 ]. In Mexico, there are also differences in socio-economic conditions resulting from geographic area and political organization. These may affect various health parameters, including mortality, morbidity and reproductive behavior. Several studies have shown different results on whether socioeconomic factors affect pregnancy outcomes and newborn conditions [ 19 - 21 ]. The inconsistency of these findings may be due to poor clarification of the mechanisms by which socioeconomic status affects LBW. This is especially true in relationship to the mother's nutritional conditions, although low maternal weight before pregnancy and small weight gain during pregnancy have been shown to be associated with higher risks of preterm infants and LBW [ 22 - 25 ]. Other studies have questioned whether maternal occupation or educational level is associated with LBW, or whether the latter is related to a group of socioeconomic factors. In this study, we have analyzed the socio-economic factors related to birthweight adjusted by others known factors in the urban population of Mexico in three hospitals located in three different geographic areas of Mexico City.
Background
On average, the worldwide incidence of low birthweight (LBW) is 17% per year, making LBW an important infant health problem in many populations [ 1 ]. The incidence of LBW varies among countries, ranging from 4% to 6% in Western countries like Sweden, France, United States and Canada (UNICEF 2003) and much higher in developing countries. In Latin America, the overall LBW rate varies according to geographical region. The Pan American Health Organization has estimated that the overall LBW rate is 8.27% in South America and Mexico and ranges from 6% in Peru to 10% in Bolivia and Venezuela. In Central America the overall LBW rate is 11.2%, ranging from 6% in Belize and El Salvador to 15% in Nicaragua [ 2 ]. The LBW rate can also vary within countries. For example, in Mexico, the incidence ranges from 8.2% to 12%, depending on geographic al region [ 3 ].In addition to its impact on infant mortality, LBW has been associated with higher probabilities of infection, malnutrition and handicapping conditions during childhood, including cerebral palsy, mental deficiencies and problems related to behavior and learning during childhood [ 4 - 6 ]. Children who survive LBW have a higher incidence of diseases, retardation in cognitive development, and undernourishment. There is also evidence that LBW or its determinant factors are associated with a predisposition to higher rates of diabetes, cardiac diseases and other future chronic health problems [ 7 - 9 ].The biological processes that affect the fetus  in utero  are related to the mother's physiology, including her nutrition (mother's weight before pregnancy and history of having newborns with LBW), exercise, infections, and consumption of tobacco, alcohol and other drugs [ 10 , 11 ].in uteroDuring the fetal phase, growth depends on the nutritional condition of the mother, indicating that pregnant women should not only increase their weight but also consume essential nutrients. For many women in the developing world, however, economic, social and cultural factors make it difficult for them to obtain the necessary food and healthcare, which are closely interrelated [ 12 ].Associations between poorer child health and poverty, inequity and social exclusion have been documented worldwide and have been shown to be independent of research methods, local culture, and available health care services [ 13 ]. While the relationship between socio-economic conditions and health have been of interest and concern for centuries, recent studies have sought to identify the social factors most relevant to health [ 14 ]. Some authors consider that health therefore may be an important determinant of opportunities in life and this process, termed "selection by health", and suggests that health "selects" people in different social strata [ 15 , 16 ].Among the socio-economic factors are income, education, occupation, household leadership and gender differences related to roles within the family [ 17 , 18 ]. In Mexico, there are also differences in socio-economic conditions resulting from geographic area and political organization. These may affect various health parameters, including mortality, morbidity and reproductive behavior.Several studies have shown different results on whether socioeconomic factors affect pregnancy outcomes and newborn conditions [ 19 - 21 ]. The inconsistency of these findings may be due to poor clarification of the mechanisms by which socioeconomic status affects LBW. This is especially true in relationship to the mother's nutritional conditions, although low maternal weight before pregnancy and small weight gain during pregnancy have been shown to be associated with higher risks of preterm infants and LBW [ 22 - 25 ]. Other studies have questioned whether maternal occupation or educational level is associated with LBW, or whether the latter is related to a group of socioeconomic factors.In this study, we have analyzed the socio-economic factors related to birthweight adjusted by others known factors in the urban population of Mexico in three hospitals located in three different geographic areas of Mexico City.Methods A case-control study was carried out in three hospitals with gynaecological and obstetrics services in Mexico City during the first semester of 1996. We recruited 154 LBW newborns, defined as newborns weighing <2500 grams, from these three hospitals, as well as 474 controls chosen from births on the same day to control for time of birth. All newborns included in this study provided a statistical power of 80% when we assumed a = 0.05 (one-sided test), a case/control ratio 1:3, and a 3% difference among both groups in the low socioeconomic level [ 26 ]. Upon arrival at the hospital for delivery, women were interviewed to determine if they met our inclusion criteria. Women with a history of previous chronic conditions or those with twin or multiple pregnancies were not included. Following delivery, and after obtaining informed consent, each woman was administered a questionnaire by a trained interviewer to obtain information about socio-economic, reproductive, and nutritional factors. Socioeconomic factors included age, level of education, civil status, occupation, income and owning certain goods; reproductive factors included parity, history of preterm delivery and LBW, which were classified as positive or negative for all previous pregnancies; and nutritional factors included calcium and iron supplementation, pre-gestational weight, prenatal care, morbidity during pregnancy and tobacco exposure. Clinical records were also reviewed to verify information about each newborn. To create a socioeconomic level index, we used two variables that have been considered as proxies [ 27 ]. Ownership of goods was defined as to whether a woman owned her house or flat and if she had a car and whether the woman and her partner were employed. Using these two variables, we constructed a three category socio-economic level index, in which High indicated that the woman and her partner had jobs and goods; Medium indicated that the woman or her partner had jobs or goods; and Low indicated that the woman and/or her partner did not have jobs or goods. Statistical analysis was performed by describing sociodemographic, reproductive and prenatal care characteristics of the mother and newborn. Univariate analysis was used to evaluate the association between the independent variable (socioeconomic level) and covariates with LBW (outcome variable). Stratified Mantel-Haenszel analysis was performed to evaluate confounding and/or interaction (i.e. parity and age). To obtain the association magnitude of the socioeconomic level adjusted by the covariates, multivariate analysis was performed (logistic regression). Variables with clinical and statistical significance were included in multivariate modeling. We used three models to classify factors relevant to LBW into three groups as known, controversial and unknown risk factors for LBW. Model 1 included socioeconomic level index and maternal age, maternal education, marital status and accessibility to public services. In model 2 we included all variables of the model 1 besides, prenatal care, tobacco exposure, and morbidity during pregnancy were considered as adjusted variables. Finally, model 3 included all variables previously mentioned (model 1 and model 2 besides reproductive variables. Analyses were performed using StataCorp. 2002. Stata Statistical Software: Release 7.0 College Station, TX:Stata Corporation.
Methods
A case-control study was carried out in three hospitals with gynaecological and obstetrics services in Mexico City during the first semester of 1996. We recruited 154 LBW newborns, defined as newborns weighing <2500 grams, from these three hospitals, as well as 474 controls chosen from births on the same day to control for time of birth. All newborns included in this study provided a statistical power of 80% when we assumed a = 0.05 (one-sided test), a case/control ratio 1:3, and a 3% difference among both groups in the low socioeconomic level [ 26 ].Upon arrival at the hospital for delivery, women were interviewed to determine if they met our inclusion criteria. Women with a history of previous chronic conditions or those with twin or multiple pregnancies were not included. Following delivery, and after obtaining informed consent, each woman was administered a questionnaire by a trained interviewer to obtain information about socio-economic, reproductive, and nutritional factors. Socioeconomic factors included age, level of education, civil status, occupation, income and owning certain goods; reproductive factors included parity, history of preterm delivery and LBW, which were classified as positive or negative for all previous pregnancies; and nutritional factors included calcium and iron supplementation, pre-gestational weight, prenatal care, morbidity during pregnancy and tobacco exposure. Clinical records were also reviewed to verify information about each newborn.To create a socioeconomic level index, we used two variables that have been considered as proxies [ 27 ]. Ownership of goods was defined as to whether a woman owned her house or flat and if she had a car and whether the woman and her partner were employed. Using these two variables, we constructed a three category socio-economic level index, in which High indicated that the woman and her partner had jobs and goods; Medium indicated that the woman or her partner had jobs or goods; and Low indicated that the woman and/or her partner did not have jobs or goods.Statistical analysis was performed by describing sociodemographic, reproductive and prenatal care characteristics of the mother and newborn. Univariate analysis was used to evaluate the association between the independent variable (socioeconomic level) and covariates with LBW (outcome variable). Stratified Mantel-Haenszel analysis was performed to evaluate confounding and/or interaction (i.e. parity and age). To obtain the association magnitude of the socioeconomic level adjusted by the covariates, multivariate analysis was performed (logistic regression). Variables with clinical and statistical significance were included in multivariate modeling. We used three models to classify factors relevant to LBW into three groups as known, controversial and unknown risk factors for LBW.Model 1 included socioeconomic level index and maternal age, maternal education, marital status and accessibility to public services. In model 2 we included all variables of the model 1 besides, prenatal care, tobacco exposure, and morbidity during pregnancy were considered as adjusted variables. Finally, model 3 included all variables previously mentioned (model 1 and model 2 besides reproductive variables.Analyses were performed using StataCorp. 2002. Stata Statistical Software: Release 7.0 College Station, TX:Stata Corporation.Results The sociodemographic characteristics of the LBW and normal birthweight groups are shown in Table  1 . There was a high proportion of married women in both groups, and 90% of women in both groups had access to public services, including electricity, water and a sewage system. The effect of maternal education had the expected direction, although it was not statistically significant. Women in the lower socio-economic level had a higher risk for LBW (OR, 2.19; 95% CI, 1.18–4.07) than those in the medium and high socioeconomic levels. Table 1 Sociodemographic characteristics of mothers of low and normal birthweight infants Variable Birthweight OR 95% CI <2500 grs ( n = 158 ) ≥ 2500 grs ( n = 474 ) Socioeconomic level    High 81 267 1.0    Medium 57 177 1.06 [0.71,1.56]    Low 20 30 2.19* [1.18,4.07] Maternal age (years)    <19 22 83 0.77 [0.46,1.29]    20–30 110 321 1.0    >30 26 70 1.08 [0.65,1.80] Maternal education (years)    >12 13 50 1.0    10–12 56 143 1.50 [0.76,2.98]    7–9 52 165 1.21 [0.61,2.40]    <7 37 116 1.22 [0.60,2.50] Marital Status    Married 138 429 1.0    Unmarried 20 45 1.38 [0.78,2.42] Accessibility to public services    No 12 51 0.68 [0.35,1.31]    Yes 146 423 1.0 * p < 0.05 We observed no significant between group differences in tobacco exposure prior to or during pregnancy, or in the frequency of urinary tract infections (Table  2 ). However hypertension (OR, 1.53; 95% CI, 0.93–2.53) and calcium supplementation (OR, 1.86; 95% CI, 0.97–3.56) during pregnancy were marginally significant. Table 2 Tobacco exposure, morbidity during pregnancy and prenatal care characteristics of mothers of low and normal birthweight infants Variable Birthweight OR 95% CI <2500 grs ( n = 158 ) ≥ 2500 grs ( n = 474 ) Smoking before pregnancy    No 126 384 1.0    Yes 32 90 1.09 [0.70,1.71] Smoking during pregnancy    No 149 451 1.0    Yes 9 23 1.19 [0.53,2.66] Hypertension during pregnancy    No 131 416 1.0    Yes 27 58 1.53 [0.93,2.53] Urinary tract infection during pregnancy    No 100 318 1.0    Yes 58 156 1.22 [0.83,1.77] Calcium supplementation    Yes 146 409 1.0    No 12 65 1.86 [0.97,3.56] Iron supplementation    Yes 64 216 1.0    No 94 258 1.20 [0.83,1.74] Prenatal care    Yes 150 454 1.0    No 8 20 1.07 [0.45,2.52] Reproductive characteristics, including maternal weight, previous preterm-birth, and previous LBW infants, were significantly different between mothers of LBW and normal birthweight infants (Table  3 ). Table 3 Reproductive characteristicsof mothers of low and normal birthweight infants Variable Birthweight OR 95% CI <2500 grs ( n = 158 ) ≥ 2500 grs ( n = 474 ) Parity   Multiparous 92 284 1.0   Primiparous 66 190 1.04 [0.73,1.51] Maternal weight (kgs)    <48 54 107 1.61* [1.09,3.84]    49–54 37 120 0.99 [0.58,1.67]    55–60 37 119 1.0    >60 30 128 0.75 [0.43,1.29] Previous a  preterm birth    No 58 213 1.0    Yes 34 71 1.75* [1.06,2.90] Previous low birth weight a    No 66 246 1.0    Yes 26 38 2.55* [1.44,4.50] Previous abortion a    No 67 222 1.0    Yes 25 62 1.33 [0.77,2.28] *p <0.05 a Only applicable for women with more than one pregnancy Multivariate analysis showed that low socioeconomic level was the most important risk factor for LBW (Table  4 ). Using model 1, the OR was 2.57 (95% CI, 1.19–5.58) after adjustments for maternal age, education, marital status, and accessibility to public services. When adjustments for prenatal care, tobacco exposure, and morbidity during pregnancy were also included (model 2), we obtained an OR of 2.45 (95% CI, 1.13–5.36) for low socioeconomic level as a risk factor for LBW (Table  4 ). When we also included adjustments for reproductive variables (model 3), we obtained an OR of 2.68 (95% CI, 1.19–6.03) for low socioeconomic level as a risk factor for LBW (Table  4 ). Table 4 Adjusted odds ratios for LBW infants by socioeconomic level among women of Mexico City Variable Model 1 Model 2 Model 3 OR 95% CI OR 95% CI OR 95% CI Socioeconomic level Medium 1.14 [0.75,1.74] 1.16 [0.76,1.77] 1.12 [0.72,1.74] Low 2.57 [1.19,5.58] 2.45 [1.13,5.36] 2.68 [1.19,6.03] Maternal age (years) <19 0.74 [0.43,1.25] 0.74 [0.43,1.26] 0.57 [0.32,1.03] >30 1.17 [0.70,1.98] 1.09 [0.64,1.84] 1.34 [0.76,2.38] Maternal education (years) < 7 1.21 [0.58,2.56] 1.11 [0.52,2.40] 1.05 [0.47,2.34] 7–9 1.29 [0.63,2.62] 1.23 [0.59,2.55] 1.39 [0.66,2.94] 10–12 1.56 [0.78,3.13] 1.61 [0.79,3.27] 1.82 [0.87,3.77] Unmarried 0.99 [0.49,2.01] 1.08 [0.52,2.28] 0.96 [0.44,2.08] No accessibility to public services 0.68 [0.34,1.33] 0.58 [0.29,1.17] 0.56 [0.27,1.14] Smoking before pregnancy 0.99 [0.57,1.72] 1.07 [0.60,1.90] Smoking during pregnancy 1.09 [0.43,2.78] 1.14 [0.43,2.99] Hypertension 1.61 [0.95,2.73] 1.55 [0.90,2.68] Urinary tract infection 1.23 [0.83,1.83] 1.14 [0.76,1.71] No calcium supplementation 1.98 [1.01,3.87] 2.30 [1.14,4.63] No iron supplementation 1.23 [0.84,1.80] 1.32 [0.89,1.96] No prenatal care 1.15 [0.47,2.87] 1.27 [0.52,3.19] Primiparous 1.73 [1.05,2.83] Pre-gestational weight(kgs) <48 2.33 [1.33,4.08] 49–54 1.28 [0.71,2.31] >60 1.32 [0.74,2.36] Previous preterm birth a 2.95 [1.04,8.38]** Previous low birth weight a 2.61 [1.36,5.04] Previous abortion a 0.52 [0.17,1.63] a only applicable for women with more than one pregnancy.
Results
The sociodemographic characteristics of the LBW and normal birthweight groups are shown in Table  1 . There was a high proportion of married women in both groups, and 90% of women in both groups had access to public services, including electricity, water and a sewage system. The effect of maternal education had the expected direction, although it was not statistically significant. Women in the lower socio-economic level had a higher risk for LBW (OR, 2.19; 95% CI, 1.18–4.07) than those in the medium and high socioeconomic levels.Sociodemographic characteristics of mothers of low and normal birthweight infants<2500 grs ( n = 158 )n = 158≥ 2500 grs ( n = 474 )n = 474HighMediumLow<1920–30>30>1210–127–9<7MarriedUnmarriedNoYes* p < 0.05We observed no significant between group differences in tobacco exposure prior to or during pregnancy, or in the frequency of urinary tract infections (Table  2 ). However hypertension (OR, 1.53; 95% CI, 0.93–2.53) and calcium supplementation (OR, 1.86; 95% CI, 0.97–3.56) during pregnancy were marginally significant.Tobacco exposure, morbidity during pregnancy and prenatal care characteristics of mothers of low and normal birthweight infants<2500 grs ( n = 158 )n = 158≥ 2500 grs ( n = 474 )n = 474NoYesNoYesNoYesNoYesYesNoYesNoYesNoReproductive characteristics, including maternal weight, previous preterm-birth, and previous LBW infants, were significantly different between mothers of LBW and normal birthweight infants (Table  3 ).Reproductive characteristicsof mothers of low and normal birthweight infants<2500 grs ( n = 158 )n = 158≥ 2500 grs ( n = 474 )n = 474MultiparousPrimiparous<4849–5455–60>60NoYesNoYesNoYes*p <0.05a Only applicable for women with more than one pregnancyMultivariate analysis showed that low socioeconomic level was the most important risk factor for LBW (Table  4 ). Using model 1, the OR was 2.57 (95% CI, 1.19–5.58) after adjustments for maternal age, education, marital status, and accessibility to public services. When adjustments for prenatal care, tobacco exposure, and morbidity during pregnancy were also included (model 2), we obtained an OR of 2.45 (95% CI, 1.13–5.36) for low socioeconomic level as a risk factor for LBW (Table  4 ). When we also included adjustments for reproductive variables (model 3), we obtained an OR of 2.68 (95% CI, 1.19–6.03) for low socioeconomic level as a risk factor for LBW (Table  4 ).Adjusted odds ratios for LBW infants by socioeconomic level among women of Mexico CityVariableModel 1Model 2Model 3Socioeconomic levelMaternal age (years)Maternal education (years)UnmarriedNo accessibility to public servicesSmoking before pregnancySmoking during pregnancyHypertensionUrinary tract infectionNo calcium supplementationNo iron supplementationNo prenatal carePrimiparousPre-gestational weight(kgs)Previous preterm birth aPrevious low birth weight aPrevious abortion aa only applicable for women with more than one pregnancy.Discussion LBW is a public health problem linked to lack of equity in populations. Despite efforts to decrease the proportion of newborns with LBW, success has been quite limited, and the problem persists in both developing and developed countries [ 28 ]. In recent years, studies focused on explaining how social factors influence this problem have shown that populations with greater inequities have a greater proportion of newborns with LBW [ 25 ]. These inequities are caused by both social conditions of populations and gender differences. Although these differences are not explicit and conclusive, they are revealed by social indicators such as access to health care services, occupation, income, education, and social exclusion or isolation. Differences found in the results about the effect of socioeconomic factors on LBW are probably due to the use of different socioeconomic indicators. It should be pointed out, however, that obtaining information that accurately reflects social and economic characteristics can be difficult, leading to the generation of proxy variables. Thus, education has been used as a proxy variable of social class, and occupation has been used as a proxy of socioeconomic status [ 29 ]. In addition, studies performed in European countries have used education as a proxy for socioeconomic level [ 29 , 30 ]. A recent Mexican study, however, found no association between LBW and household infrastructure (including lack of indoor sanitation or water facilities and lack of electricity) when used as a socioeconomic indicator, although this study found that LBW was related mainly to access or utilization of prenatal services and disadvantageous maternal lifestyle behaviors [ 31 ]. In our study, we used ownership of goods and having a job as indicators of socioeconomic level in a Mexican urban population with social security. Using these indicators, we found that low socioeconomic level is the most important risk factor for LBW, independent of other factors such as reproductive and nutritional characteristics, smoking, morbidity during pregnancy, and accessibility to health services and prenatal care. These results are similar to those of other studies describing a positive relationship between socioeconomic condition and effects on health [ 32 - 34 ] It is important to mention that to diminish the possible error of misclassifying we used at least two indicators to construct the socioeconomic level variable (ownership of goods and occupation). However we thought that it will be a non differential error. Although we analyzed other proxies of socioeconomic level, we found that each of them was irrelevant. For example, although education has been used as a proxy of social status, we found that this indicator was not important in our population, perhaps because the women in our study were incorporated into the work force, making exposure to intermediate factors, such as occupational stress and load work, an influence on pregnancy outcomes. Family circumstances and biological processes may also be affected by a wider social context, including cultural and historic issues such as educational opportunities, parent's divorce, unemployment, risk of poverty risk, and risks factors for smoking and obesity. Although many socioeconomic factors related to LBW have been identified, the specific role of each of them is not known, limiting the ability to use preventive actions in exposed populations. Interventions aimed at reducing the number of LBW infants have had limited success on conditions of the newborn, although some showed benefits in pregnant women [ 35 ]. To decrease the incidence of LBW, it is important to consider health services interventions to get better quality of care for pregnant women. Investments in the health and development of the most vulnerable populations, such as pregnant women and children, are important in themselves, because they prepare the context and environment for a more productive and healthy life, with the full development and use of mental and physical human potential [ 36 ].
Discussion
LBW is a public health problem linked to lack of equity in populations. Despite efforts to decrease the proportion of newborns with LBW, success has been quite limited, and the problem persists in both developing and developed countries [ 28 ]. In recent years, studies focused on explaining how social factors influence this problem have shown that populations with greater inequities have a greater proportion of newborns with LBW [ 25 ]. These inequities are caused by both social conditions of populations and gender differences. Although these differences are not explicit and conclusive, they are revealed by social indicators such as access to health care services, occupation, income, education, and social exclusion or isolation.Differences found in the results about the effect of socioeconomic factors on LBW are probably due to the use of different socioeconomic indicators. It should be pointed out, however, that obtaining information that accurately reflects social and economic characteristics can be difficult, leading to the generation of proxy variables. Thus, education has been used as a proxy variable of social class, and occupation has been used as a proxy of socioeconomic status [ 29 ]. In addition, studies performed in European countries have used education as a proxy for socioeconomic level [ 29 , 30 ]. A recent Mexican study, however, found no association between LBW and household infrastructure (including lack of indoor sanitation or water facilities and lack of electricity) when used as a socioeconomic indicator, although this study found that LBW was related mainly to access or utilization of prenatal services and disadvantageous maternal lifestyle behaviors [ 31 ].In our study, we used ownership of goods and having a job as indicators of socioeconomic level in a Mexican urban population with social security. Using these indicators, we found that low socioeconomic level is the most important risk factor for LBW, independent of other factors such as reproductive and nutritional characteristics, smoking, morbidity during pregnancy, and accessibility to health services and prenatal care. These results are similar to those of other studies describing a positive relationship between socioeconomic condition and effects on health [ 32 - 34 ]It is important to mention that to diminish the possible error of misclassifying we used at least two indicators to construct the socioeconomic level variable (ownership of goods and occupation). However we thought that it will be a non differential error. Although we analyzed other proxies of socioeconomic level, we found that each of them was irrelevant. For example, although education has been used as a proxy of social status, we found that this indicator was not important in our population, perhaps because the women in our study were incorporated into the work force, making exposure to intermediate factors, such as occupational stress and load work, an influence on pregnancy outcomes. Family circumstances and biological processes may also be affected by a wider social context, including cultural and historic issues such as educational opportunities, parent's divorce, unemployment, risk of poverty risk, and risks factors for smoking and obesity.Although many socioeconomic factors related to LBW have been identified, the specific role of each of them is not known, limiting the ability to use preventive actions in exposed populations. Interventions aimed at reducing the number of LBW infants have had limited success on conditions of the newborn, although some showed benefits in pregnant women [ 35 ]. To decrease the incidence of LBW, it is important to consider health services interventions to get better quality of care for pregnant women. Investments in the health and development of the most vulnerable populations, such as pregnant women and children, are important in themselves, because they prepare the context and environment for a more productive and healthy life, with the full development and use of mental and physical human potential [ 36 ].Conclusion Heterogeneity among different populations makes findings related to interventions in one population not applicable to others. Thus, it is necessary to design studies that account for the geographic, racial, cultural, social and economic context of each country and specific group. In the case of Mexico, we believe that our results provide information representing a closer approach to the effect of socioeconomic status on LBW. Our results provide a starting point in the search for better indicators for evaluating socioeconomic status in Mexican populations with other social conditions, including suburban, rural, and indigenous populations. Another aspect to consider within the context of each country is the availability of information. In Mexico, there is no way to obtain uniform information about the socioeconomic characteristics of the population. Information about conditions of different populations is important in designing programs aimed at solving existing inequities. It is a challenge to create a socioeconomic index that will reflect the real living conditions of pregnant women in Mexico. This effort will require including other variables, as well as refining proxy variables.
Conclusion
Heterogeneity among different populations makes findings related to interventions in one population not applicable to others. Thus, it is necessary to design studies that account for the geographic, racial, cultural, social and economic context of each country and specific group. In the case of Mexico, we believe that our results provide information representing a closer approach to the effect of socioeconomic status on LBW. Our results provide a starting point in the search for better indicators for evaluating socioeconomic status in Mexican populations with other social conditions, including suburban, rural, and indigenous populations.Another aspect to consider within the context of each country is the availability of information. In Mexico, there is no way to obtain uniform information about the socioeconomic characteristics of the population. Information about conditions of different populations is important in designing programs aimed at solving existing inequities. It is a challenge to create a socioeconomic index that will reflect the real living conditions of pregnant women in Mexico. This effort will require including other variables, as well as refining proxy variables.Competing interests The author(s) declare that they have no competing interests.
Competing interests
The author(s) declare that they have no competing interests.Authors' contributions LPTA, PCC, SFH contributed in the conception and design of the study and statistical analysis. JPVB, ERM reviewed for important intellectual content. All authors participated in the interpretation of data and read and approved the final version to be published.
Authors' contributions
LPTA, PCC, SFH contributed in the conception and design of the study and statistical analysis. JPVB, ERM reviewed for important intellectual content. All authors participated in the interpretation of data and read and approved the final version to be published.Pre-publication history The pre-publication history for this paper can be accessed here:
Pre-publication history
The pre-publication history for this paper can be accessed here:


Introduction The C677T MTHFR variant has been associated with the same third trimester pregnancy complications as seen in women who have elevations of maternal serum α-fetoprotein (MSAFP). We hypothesized that these women with third trimester pregnancy complications and MSAFP elevations would have an increased frequency of the variant compared to an abnormal study control group (women with MSAFP elevations without pregnancy complications) as well as to normal population controls.
Introduction
The C677T MTHFR variant has been associated with the same third trimester pregnancy complications as seen in women who have elevations of maternal serum α-fetoprotein (MSAFP). We hypothesized that these women with third trimester pregnancy complications and MSAFP elevations would have an increased frequency of the variant compared to an abnormal study control group (women with MSAFP elevations without pregnancy complications) as well as to normal population controls.Methods Women who had unexplained elevations of MSAFP in pregnancy were ascertained retrospectively. The frequency of the C677T MTHFR variant among those women with unexplained elevations of MSAFP who had experienced later pregnancy complications was compared to that of women with unexplained elevations of MSAFP without complications as well as to that of the previously established Manitoba frequency.
Methods
Women who had unexplained elevations of MSAFP in pregnancy were ascertained retrospectively. The frequency of the C677T MTHFR variant among those women with unexplained elevations of MSAFP who had experienced later pregnancy complications was compared to that of women with unexplained elevations of MSAFP without complications as well as to that of the previously established Manitoba frequency.Results Women who had complications of pregnancy and an unexplained MSAFP elevation had a higher allele frequency for the C677T MTHFR variant (q = 0.36,) compared to women with MSAFP elevations and normal pregnancy outcomes (q = 0.25, OR 1.73 95% CI 1.25–2.37,  p  = 0.03). The frequency was also higher than that of the population controls (q= 0.25, OR 1.70 95% CI 1.11–2.60,  p  = 0.007). The frequency in women with MSAFP elevations without pregnancy complications was not significantly different from that of the population controls ( p  = 0.41).
Results
Women who had complications of pregnancy and an unexplained MSAFP elevation had a higher allele frequency for the C677T MTHFR variant (q = 0.36,) compared to women with MSAFP elevations and normal pregnancy outcomes (q = 0.25, OR 1.73 95% CI 1.25–2.37,  p  = 0.03). The frequency was also higher than that of the population controls (q= 0.25, OR 1.70 95% CI 1.11–2.60,  p  = 0.007). The frequency in women with MSAFP elevations without pregnancy complications was not significantly different from that of the population controls ( p  = 0.41).pppConclusion Women with unexplained elevations of MSAFP and who experience complications in later pregnancy are more likely to have one or two alleles of the C677T MTHFR variant.
Conclusion
Women with unexplained elevations of MSAFP and who experience complications in later pregnancy are more likely to have one or two alleles of the C677T MTHFR variant.

Background Significant elevations of amniotic fluid and maternal serum alpha-fetoprotein (MSAFP) have been shown to be associated with spina bifida and other neural tube defects (NTD). The province of Manitoba, Canada offers province-wide midtrimester MSAFP screening to all pregnant women. It has been recognized that some pregnant women with midtrimester unexplained elevations of MSAFP [ 1 , 2 ]. Increased total plasma homocysteine alters placental function and has been associated with the same complications that are associated with unexplained elevated MSAFP [ 3 - 7 ]. The C677T MTHFR variant has also been associated with complications of pregnancy in some, but not all, studies [ 8 - 10 ] C677T MTHFR may therefore contribute to complications of pregnancy by elevating serum homocysteine. Poor placental function could result in both an unexplained elevation of MSAFP and complications of pregnancy. We therefore hypothesized that women with third trimester pregnancy complications and MSAFP elevations (cases) would have an increased frequency of the variant compared to the Manitoba population (population controls) or women with MSAFP elevations without pregnancy complications (study controls) if they had low folate intake.
Background
Significant elevations of amniotic fluid and maternal serum alpha-fetoprotein (MSAFP) have been shown to be associated with spina bifida and other neural tube defects (NTD). The province of Manitoba, Canada offers province-wide midtrimester MSAFP screening to all pregnant women. It has been recognized that some pregnant women with midtrimester unexplained elevations of MSAFP [ 1 , 2 ].Increased total plasma homocysteine alters placental function and has been associated with the same complications that are associated with unexplained elevated MSAFP [ 3 - 7 ]. The C677T MTHFR variant has also been associated with complications of pregnancy in some, but not all, studies [ 8 - 10 ] C677T MTHFR may therefore contribute to complications of pregnancy by elevating serum homocysteine. Poor placental function could result in both an unexplained elevation of MSAFP and complications of pregnancy. We therefore hypothesized that women with third trimester pregnancy complications and MSAFP elevations (cases) would have an increased frequency of the variant compared to the Manitoba population (population controls) or women with MSAFP elevations without pregnancy complications (study controls) if they had low folate intake.Methods Background to methods In a small pilot study of 32 couples, we found that women who had an unexplained elevation of MSAFP and a normal midtrimester fetal ultrasound, and their partners, had a significantly increased C677T MTHFR frequency compared to Manitoba newborns (RR 1.42, 95% CI 1.08–1.85, p = 0.012, two tailed) [ 11 ]. The newborn study that examined 977 anonymous consecutive neonatal screening blood spots showed that 36% of Manitoba newborns were heterozygous and 7% were homozygous for C677T MTHFR [ 12 ] (q = 0.25). Subsequently, on evaluation of the pregnancy outcomes of our pilot study women, we noted that, among eight women who had gone on to experience complications of pregnancy, the odds ratio for having the C677T MTHFR allele was 2.3 times higher than in the Manitoba population. However, the result was not statistically significant ( p  = 0.151, two tailed) indicating the frequency was increased but, this could have been a random result. Ascertainment and recruitment of study population All pregnant women in Manitoba are eligible for routine serum screening through the voluntary MMSSP. In Manitoba, an elevation of MSAFP is defined as 2.3 multiples of the median (MOM) or greater. Candidates for inclusion in this study were women with an unexplained MSAFP elevation (i.e. not due to fetal anomalies, incorrect estimation of gestational age, previously unrecognized fetal demise, or multiple gestation) with either a complicated or uncomplicated pregnancy outcome. After appropriate approvals had been obtained from The University of Manitoba Health Research Ethics Board, review of the screening records began in 1999 and took three years. For a study using a two step consent to participate methodology administered by mail, the expected response rate (after excluding lost to follow-up) would be 20% [ 13 ]. Our goal was 1000 invitations. We anticipated this would result in approximately 120 participants. This would be double the minimum number of participants suggested by the power analysis we had conducted for the pilot study. To increase our response rate further, we added telephone follow-up for invited potential participants who were non-responders [ 14 ]. All screening records from 1995–1999 were reviewed, accounting for 783 invitations. Records for 2000–2002 were reviewed systematically as outcome information on each pregnancy became available to MMSSP. Records for 1990–1994 were then reviewed systematically in order to bring the total up to 1000. If a woman had more than one pregnancy with an elevation of MSAFP screened by the MMSSP, only the first pregnancy encountered in the retrospective review was used for the study. Previous or subsequent pregnancies were not included. Women with preexisting conditions known to influence pregnancy outcome, such as essential hypertension, and mothers of babies with major congenital anomalies were excluded. Eight women who had relinquished their babies for adoption or whose babies were placed in foster care were also excluded. Women who met the inclusion criteria were divided into two groups for analysis. Cases were defined as women with pregnancies complicated by one of the complications previously shown to be associated with an unexplained elevation of MSAFP at midtrimester [ 2 ]. These include: intrauterine growth restriction (IUGR) (<10 th  percentile), pregnancy induced hypertension, preeclampsia, eclampsia, postpartum hemorrhage, retained placenta requiring manual delivery, abruptio placenta, premature delivery (<36 weeks gestation or requiring specialized neonatal care for prematurity) and unexplained fetal demise. Study controls were women with normal outcomes which were defined as those with delivery at term ≥ 36 weeks gestation), no complications of pregnancy, a normal placenta and a healthy baby. Definition of complications was based on ICDC-9 codes in the MMSSP outcome charts for each patient [ 15 ] which are then confirmed later by chart review for all those with a positive MMSSP result. All women ascertained as having unexplained MSAFP elevations and who fit the inclusion criteria above, were invited by letter to participate. The previously reported newborn study provided population control group data [ 12 ]. Study questionnaires Women who agreed to participate in the study were mailed the appropriate questionnaires and blood requisitions. The questionnaire included a semi-quantitative food frequency questionnaire (FFQ) based on standard methodology but, modified to suit Manitoba residents and previously validated for this population by biochemical analysis during the pilot study [ 11 , 16 ]. The survey included questions on vitamin supplement intake to determine preconceptional or prenatal supplementation as well as current use of vitamins. Dietary intake of folate and folic acid from supplements, and intake of the cofactors B 12  and B 6 , were calculated from the FFQ for intake both during pregnancy and at the time of the study. A correction of an additional 0.1 mg for folic acid fortification that began in Canada in 1998 was included for pregnancies that began after fortification [ 17 ]. FFQ analyses were performed with the researcher blinded as to the status of the mother. Laboratory analysis Total plasma homocysteine, red blood cell folate, and serum folate were determined using established methodology [ 18 , 19 ]. Samples were processed on site with clotting and separation by spinning. Sera was stored at 4°C during shipping to the central laboratory and until processing. DNA was extracted from whole blood and C677T MTHFR genotyping was performed using previously established methodology [ 11 , 20 , 21 ]. Genotyping and biochemical analyses were performed also blinded. Statistical analysis Chi-squared analysis (one tailed unless otherwise noted) was used for allele frequency. Comparisons of potentially confounding factors between the case group and the study control group were undertaken. Parametric data were analyzed with the Student's  t  test for difference between means with Bonferroni correction for multiple comparisons. Data not normally distributed were analyzed using the nonparametric Mann-Whitney Rank Sum Test. Linear regression was used to test the validity of the dietary survey. A multivariate analysis included age, smoking, maternal weight at the time of MSAFP testing, presence of C677T MTHFR, gender and weight of infant, biochemical parameters, and FFQ results for folate, B 12  and B 6 , both at the time of the survey and for during the pregnancy was undertaken. In order to avoid convergence due to the large number of variables, the analysis was completed in subsets of six variables. Variables with the higher association scores from these analyses were then combined for further testing in various combinations using stepwise multiple linear regression. Also linear regression analysis of each continuous variable with genotype results was performed. Corrections for multiple comparisons were included. Software used was NCSS Statistical Systems for Windows [ 22 ].
Methods
Background to methods In a small pilot study of 32 couples, we found that women who had an unexplained elevation of MSAFP and a normal midtrimester fetal ultrasound, and their partners, had a significantly increased C677T MTHFR frequency compared to Manitoba newborns (RR 1.42, 95% CI 1.08–1.85, p = 0.012, two tailed) [ 11 ]. The newborn study that examined 977 anonymous consecutive neonatal screening blood spots showed that 36% of Manitoba newborns were heterozygous and 7% were homozygous for C677T MTHFR [ 12 ] (q = 0.25). Subsequently, on evaluation of the pregnancy outcomes of our pilot study women, we noted that, among eight women who had gone on to experience complications of pregnancy, the odds ratio for having the C677T MTHFR allele was 2.3 times higher than in the Manitoba population. However, the result was not statistically significant ( p  = 0.151, two tailed) indicating the frequency was increased but, this could have been a random result.
Background to methods
In a small pilot study of 32 couples, we found that women who had an unexplained elevation of MSAFP and a normal midtrimester fetal ultrasound, and their partners, had a significantly increased C677T MTHFR frequency compared to Manitoba newborns (RR 1.42, 95% CI 1.08–1.85, p = 0.012, two tailed) [ 11 ]. The newborn study that examined 977 anonymous consecutive neonatal screening blood spots showed that 36% of Manitoba newborns were heterozygous and 7% were homozygous for C677T MTHFR [ 12 ] (q = 0.25). Subsequently, on evaluation of the pregnancy outcomes of our pilot study women, we noted that, among eight women who had gone on to experience complications of pregnancy, the odds ratio for having the C677T MTHFR allele was 2.3 times higher than in the Manitoba population. However, the result was not statistically significant ( p  = 0.151, two tailed) indicating the frequency was increased but, this could have been a random result.pAscertainment and recruitment of study population All pregnant women in Manitoba are eligible for routine serum screening through the voluntary MMSSP. In Manitoba, an elevation of MSAFP is defined as 2.3 multiples of the median (MOM) or greater. Candidates for inclusion in this study were women with an unexplained MSAFP elevation (i.e. not due to fetal anomalies, incorrect estimation of gestational age, previously unrecognized fetal demise, or multiple gestation) with either a complicated or uncomplicated pregnancy outcome. After appropriate approvals had been obtained from The University of Manitoba Health Research Ethics Board, review of the screening records began in 1999 and took three years. For a study using a two step consent to participate methodology administered by mail, the expected response rate (after excluding lost to follow-up) would be 20% [ 13 ]. Our goal was 1000 invitations. We anticipated this would result in approximately 120 participants. This would be double the minimum number of participants suggested by the power analysis we had conducted for the pilot study. To increase our response rate further, we added telephone follow-up for invited potential participants who were non-responders [ 14 ]. All screening records from 1995–1999 were reviewed, accounting for 783 invitations. Records for 2000–2002 were reviewed systematically as outcome information on each pregnancy became available to MMSSP. Records for 1990–1994 were then reviewed systematically in order to bring the total up to 1000. If a woman had more than one pregnancy with an elevation of MSAFP screened by the MMSSP, only the first pregnancy encountered in the retrospective review was used for the study. Previous or subsequent pregnancies were not included. Women with preexisting conditions known to influence pregnancy outcome, such as essential hypertension, and mothers of babies with major congenital anomalies were excluded. Eight women who had relinquished their babies for adoption or whose babies were placed in foster care were also excluded. Women who met the inclusion criteria were divided into two groups for analysis. Cases were defined as women with pregnancies complicated by one of the complications previously shown to be associated with an unexplained elevation of MSAFP at midtrimester [ 2 ]. These include: intrauterine growth restriction (IUGR) (<10 th  percentile), pregnancy induced hypertension, preeclampsia, eclampsia, postpartum hemorrhage, retained placenta requiring manual delivery, abruptio placenta, premature delivery (<36 weeks gestation or requiring specialized neonatal care for prematurity) and unexplained fetal demise. Study controls were women with normal outcomes which were defined as those with delivery at term ≥ 36 weeks gestation), no complications of pregnancy, a normal placenta and a healthy baby. Definition of complications was based on ICDC-9 codes in the MMSSP outcome charts for each patient [ 15 ] which are then confirmed later by chart review for all those with a positive MMSSP result. All women ascertained as having unexplained MSAFP elevations and who fit the inclusion criteria above, were invited by letter to participate. The previously reported newborn study provided population control group data [ 12 ].
Ascertainment and recruitment of study population
All pregnant women in Manitoba are eligible for routine serum screening through the voluntary MMSSP. In Manitoba, an elevation of MSAFP is defined as 2.3 multiples of the median (MOM) or greater. Candidates for inclusion in this study were women with an unexplained MSAFP elevation (i.e. not due to fetal anomalies, incorrect estimation of gestational age, previously unrecognized fetal demise, or multiple gestation) with either a complicated or uncomplicated pregnancy outcome. After appropriate approvals had been obtained from The University of Manitoba Health Research Ethics Board, review of the screening records began in 1999 and took three years. For a study using a two step consent to participate methodology administered by mail, the expected response rate (after excluding lost to follow-up) would be 20% [ 13 ]. Our goal was 1000 invitations. We anticipated this would result in approximately 120 participants. This would be double the minimum number of participants suggested by the power analysis we had conducted for the pilot study. To increase our response rate further, we added telephone follow-up for invited potential participants who were non-responders [ 14 ].All screening records from 1995–1999 were reviewed, accounting for 783 invitations. Records for 2000–2002 were reviewed systematically as outcome information on each pregnancy became available to MMSSP. Records for 1990–1994 were then reviewed systematically in order to bring the total up to 1000. If a woman had more than one pregnancy with an elevation of MSAFP screened by the MMSSP, only the first pregnancy encountered in the retrospective review was used for the study. Previous or subsequent pregnancies were not included. Women with preexisting conditions known to influence pregnancy outcome, such as essential hypertension, and mothers of babies with major congenital anomalies were excluded. Eight women who had relinquished their babies for adoption or whose babies were placed in foster care were also excluded.Women who met the inclusion criteria were divided into two groups for analysis. Cases were defined as women with pregnancies complicated by one of the complications previously shown to be associated with an unexplained elevation of MSAFP at midtrimester [ 2 ]. These include: intrauterine growth restriction (IUGR) (<10 th  percentile), pregnancy induced hypertension, preeclampsia, eclampsia, postpartum hemorrhage, retained placenta requiring manual delivery, abruptio placenta, premature delivery (<36 weeks gestation or requiring specialized neonatal care for prematurity) and unexplained fetal demise. Study controls were women with normal outcomes which were defined as those with delivery at term ≥ 36 weeks gestation), no complications of pregnancy, a normal placenta and a healthy baby. Definition of complications was based on ICDC-9 codes in the MMSSP outcome charts for each patient [ 15 ] which are then confirmed later by chart review for all those with a positive MMSSP result. All women ascertained as having unexplained MSAFP elevations and who fit the inclusion criteria above, were invited by letter to participate. The previously reported newborn study provided population control group data [ 12 ].Study questionnaires Women who agreed to participate in the study were mailed the appropriate questionnaires and blood requisitions. The questionnaire included a semi-quantitative food frequency questionnaire (FFQ) based on standard methodology but, modified to suit Manitoba residents and previously validated for this population by biochemical analysis during the pilot study [ 11 , 16 ]. The survey included questions on vitamin supplement intake to determine preconceptional or prenatal supplementation as well as current use of vitamins. Dietary intake of folate and folic acid from supplements, and intake of the cofactors B 12  and B 6 , were calculated from the FFQ for intake both during pregnancy and at the time of the study. A correction of an additional 0.1 mg for folic acid fortification that began in Canada in 1998 was included for pregnancies that began after fortification [ 17 ]. FFQ analyses were performed with the researcher blinded as to the status of the mother.
Study questionnaires
Women who agreed to participate in the study were mailed the appropriate questionnaires and blood requisitions. The questionnaire included a semi-quantitative food frequency questionnaire (FFQ) based on standard methodology but, modified to suit Manitoba residents and previously validated for this population by biochemical analysis during the pilot study [ 11 , 16 ]. The survey included questions on vitamin supplement intake to determine preconceptional or prenatal supplementation as well as current use of vitamins. Dietary intake of folate and folic acid from supplements, and intake of the cofactors B 12  and B 6 , were calculated from the FFQ for intake both during pregnancy and at the time of the study. A correction of an additional 0.1 mg for folic acid fortification that began in Canada in 1998 was included for pregnancies that began after fortification [ 17 ]. FFQ analyses were performed with the researcher blinded as to the status of the mother.Laboratory analysis Total plasma homocysteine, red blood cell folate, and serum folate were determined using established methodology [ 18 , 19 ]. Samples were processed on site with clotting and separation by spinning. Sera was stored at 4°C during shipping to the central laboratory and until processing. DNA was extracted from whole blood and C677T MTHFR genotyping was performed using previously established methodology [ 11 , 20 , 21 ]. Genotyping and biochemical analyses were performed also blinded.
Laboratory analysis
Total plasma homocysteine, red blood cell folate, and serum folate were determined using established methodology [ 18 , 19 ]. Samples were processed on site with clotting and separation by spinning. Sera was stored at 4°C during shipping to the central laboratory and until processing. DNA was extracted from whole blood and C677T MTHFR genotyping was performed using previously established methodology [ 11 , 20 , 21 ]. Genotyping and biochemical analyses were performed also blinded.Statistical analysis Chi-squared analysis (one tailed unless otherwise noted) was used for allele frequency. Comparisons of potentially confounding factors between the case group and the study control group were undertaken. Parametric data were analyzed with the Student's  t  test for difference between means with Bonferroni correction for multiple comparisons. Data not normally distributed were analyzed using the nonparametric Mann-Whitney Rank Sum Test. Linear regression was used to test the validity of the dietary survey. A multivariate analysis included age, smoking, maternal weight at the time of MSAFP testing, presence of C677T MTHFR, gender and weight of infant, biochemical parameters, and FFQ results for folate, B 12  and B 6 , both at the time of the survey and for during the pregnancy was undertaken. In order to avoid convergence due to the large number of variables, the analysis was completed in subsets of six variables. Variables with the higher association scores from these analyses were then combined for further testing in various combinations using stepwise multiple linear regression. Also linear regression analysis of each continuous variable with genotype results was performed. Corrections for multiple comparisons were included. Software used was NCSS Statistical Systems for Windows [ 22 ].
Statistical analysis
Chi-squared analysis (one tailed unless otherwise noted) was used for allele frequency. Comparisons of potentially confounding factors between the case group and the study control group were undertaken. Parametric data were analyzed with the Student's  t  test for difference between means with Bonferroni correction for multiple comparisons. Data not normally distributed were analyzed using the nonparametric Mann-Whitney Rank Sum Test. Linear regression was used to test the validity of the dietary survey. A multivariate analysis included age, smoking, maternal weight at the time of MSAFP testing, presence of C677T MTHFR, gender and weight of infant, biochemical parameters, and FFQ results for folate, B 12  and B 6 , both at the time of the survey and for during the pregnancy was undertaken. In order to avoid convergence due to the large number of variables, the analysis was completed in subsets of six variables. Variables with the higher association scores from these analyses were then combined for further testing in various combinations using stepwise multiple linear regression. Also linear regression analysis of each continuous variable with genotype results was performed. Corrections for multiple comparisons were included. Software used was NCSS Statistical Systems for Windows [ 22 ].tResults Participation rates Nine hundred and ninety four women were identified as eligible (342 cases and 652 controls). Of the 590 women successfully contacted, 130 (22%) agreed to participate (56 cases and 74 controls). Four hundred and four women were lost to follow-up. Cases were more likely to choose to participate than controls and this difference was significant (1.5,  p  = 0.030). There was no difference in the proportions of cases and controls that were lost to follow-up ( p  = 0.157). We had anticipated a 20% response rate and we achieved 24%. Genotype results Genotypes were available for 54 cases and 73 controls for this analysis. Results are summarized in Table  1 . The allele frequency for the C677T MTHFR variant in the Manitoba population has been previously established to be q = 0.25. Women who had complications of pregnancy and an unexplained MSAFP elevation had a higher allele frequency for the C677T MTHFR variant (q = 0.36) compared to women with MSAFP elevations and normal pregnancy outcomes (q = 0.25, OR 1.73 95% CI 1.25–2.37,  p  = 0.03). The frequency was also higher than in the population controls (q = 0.25, OR 1.70 95% CI 1.11–2.60,  p  = 0.007). The frequency in women without pregnancy complications and MSAFP elevations (study controls) was not significantly different than that seen in population controls ( p  = 0.41). Table 1 Comparison of allele frequency of C677T MTHFR between cases, study controls, and population controls. Subjects C/C (%) C/T (%) T/T (%) Comparing to study controls*  OR (95%CI) Comparing to population*  OR (95%CI) Cases N = 54 21 (39) 27 (50) 6 (11) 1.73 (1.25–2.37) ( p  = 0.033) 1.70 (1.11–2.60) ( p  = 0.007) Study Controls N = 73 40 (55) 30 (41) 3 (4) ~ 0.98 (0.46–1.55) ( p  = 0.410) Population N = 977 557 (57) 352 (36) 68 (7) 0.98 (0.46–1.55) ( p  = 0.410) ~ * χ 2  comparison of allele frequency (total T and C) in each group, one tailed. Cases: women with unexplained elevations of MSAFP who had subsequent complications of pregnancy, (C = 69, T = 39) Controls: women who had unexplained elevations of MSAFP and no subsequent complications (C = 110, T = 36) Population controls were 977 newborns (C = 1466, T = 488) [12]. C/C = normal type, C/T = heterozygous for thermolabile variant, T/T = homozygous for thermolabile variant. The case and study control groups included women at various stages of their child bearing years. Only one woman recruited as a control subject had a previous or subsequent pregnancy with an unexplained elevation of MSAFP and complications. She was a heterozygote for C677T MTHFR. No case subjects had a previous or subsequent pregnancy with an unexplained elevation of MSAFP and a normal outcome, but four case subjects had had a previous or subsequent pregnancy with complications after an elevated MSAFP. If the case versus control classification had been based on whether or not a woman had  ever  had a pregnancy with an unexplained elevation of MSAFP followed by complications, the association would still be present when compared to the population control (q = 0.3636, OR 1.72, 95%CI 1.27–2.61,  p  = 0.0055). Biochemical results Heterozygotes and homozygotes for C677T MTHFR had lower average values (r = 0.978,  p  = 0.019) for serum folate than those who did not have the variant. None of the women were deficient in either serum folate (defined as <7.0 nmol/L) or red blood cell folate (defined as <430 nmol/L RBC). There was no significant difference in mean homocysteine levels (Table  2 ). Table 2 Comparison of the parametric characteristics of women with unexplained elevations of MSAFP according to those with and without complications of pregnancy Characteristic Mean Cases (SD) Mean Controls (SD) p  value MSAFP result 2.78 (± 0.62) 3.16 (± 3.76) 0.398 weeks gestation 17.1 (± 1.61) 16.9 (± 1.40) 0.432 μg/folate/day in pregnancy 2 1216 (± 915) 1010 (± 892) 0.206 μg/folate/day at time of study 2 557 (± 341) 523 (± 498) 0.588 erc folate (nmol/L RBC) 1234 (± 289) 1208 (± 317) 0.632 serum folate (nmol/L) 32.3 (± 5.80) 32.3 (± 5.71) 0.956 serum homocysteine (μmol/L) 7.8 (± 2.26) 8.4 (± 2.80) 0.246 μg B 12 /day in pregnancy 2 12.4 (± 5.37) 13.4 (8.31) 0.488 μg B 12 /day at time of study 2 8.9 (± 12.26) 8.6 (± 10.83) 0.899 mg B 6 /day in pregnancy 2 8.4 (± 9.69) 7.2 (± 9.01) 0.461 mg B 6 /day at time of study 2 6.0 (± 13.35) 5.5 (± 11.10) 0.792 mother's age at delivery 31 (± 4.19) 30 (± 5.19) 0.251 mother's weight in Kg 76 (± 17.26) 69 (± 16.09) 0.013 1 1  This value is not significant after Bonferroni correction for multiple comparisons. See discussion.  2 Data was skewed due to a small number of women in both groups taking large dose vitamin supplements. When these women were removed from the analysis the result remained nonsignificant. Validity of surveys Seven cases and one study control declined to fill out their dietary surveys. Mean values were inserted in the multivariate analysis for these eight women. The validity of the dietary survey was demonstrated again for this study by linear regression analysis. Consistent with known homocysteine metabolism [ 23 ], a negative correlation existed between serum homocysteine and both red blood cell folate (r = 0.945  p  = 0.0052) and serum folate (r = 0.932,  p  = 0.0001). Higher intake of dietary folate (including synthetic folic acid from supplements) as reported by the FFQ for the time of study was associated with higher serum folate (r = 0.941,  p  = 0.0001) and higher red blood cell folate (r = 0.949,  p  = 0.0166). We did several checks to determine that the women were answering their surveys accurately. Comparisons of specific data items available in the women's MMSSP charts at the time of pregnancy with the data reported in the surveys showed excellent agreement for every item examined indicating women answered questions accurately. Women who reported smoking (as a quantitative value from 0–3 based on 1/2 packs/day smoked) showed a negative correlation with serum folate (r = 0.923,  p  = 0.0062) consistent with accurately reporting their smoking habits [ 24 ]. Based on the results of these tests of the validity of our surveys, we are confident that the information provided by our participants was accurate. Analysis of FFQ survey and MMSSP data The ethnicity of the infants born to the case mothers (based on the ethnicity reported for the infants grandparents) was 84% Caucasian, 5% Aboriginal. Mixed ethnicity was reported for 11% of the infants with one parent Caucasian and the other parent Aboriginal, or rarely Black or Asian. The ethnic distribution was the same for controls and is typical for the Manitoba population [ 25 , 26 ]. There were also no significant differences between cases and controls with respect to their place of residence within the province (such as rural versus urban address). There were no significant differences in dietary and supplemental intake of folate, B 12 , or B 6 , or in the biochemical parameters of case and control mothers. There was no difference in the percentage of cases and controls who reported taking prenatal vitamin supplements during pregnancy (37/48 cases and 55/72) or taking vitamin and/or folate supplements preconceptionally (17/48 cases and 25/72 study controls). We attempted to divide our cases into smaller groups by type of pregnancy complication. We also separated isolated IUGR and IUGR associated with hypertensive disorders of pregnancy. Most of the groups lacked power for statistical analysis due to small numbers. However, normotensive women whose fetus had IUGR (N = 12) had a higher frequency of the C677T MTHFR variant compared to the population controls (q = 0.33, OR 2.58 95% CI, 1.78–3.73,  p  = 0.013). Homozygosity for the C677T MTHFR variant is associated with IUGR in women who do not take vitamin supplements according to one large study of Canadian women [ 10 , 27 ]. Our findings are in agreement with this result as only 3/12 women took supplements. We found this effect in a group of combined heterozygous and homozygous women. There was a trend towards higher mean weight for mothers who had complications at the time of MSAFP test in the individual comparisons, but this was not significant after correction for multiple comparisons (Table  2  and  3 ). The multivariate analysis did not reveal any unexpected associations, but it did show the importance of maternal weight as a variable (r = 0.933,  p  = 0.024). This was also not unexpected given that some of the complications we were examining are associated with obesity [ 28 ]. Even after controlling for women's weight in the multivariate analysis, the higher frequency of C677T MTHFR among cases remained significant (r = 0.734,  p  = 0.0462). There was no association between weight and MTHFR status (r = 0.431,  p  = 0.679). Table 3 Comparison of the nonparametric characteristics of women with unexplained elevations of MSAFP according to those with and without complications of pregnancy. Nonparametric Characteristics Cases Controls p  value Location  Winnipeg 44 63 0.854  southern city 3 2  southern town 3 5  southern rural 8 8  northern city 2 1  northern town 1 4  northern rural 5 7 ethnicity 0.972  Caucasian 52 66  Aborginal 2 2  Mixed Caucasian/Aboriginal 7 6  Mixed Caucasian/Black 1 3  Asian 1 1  Mixed Caucasian/Asian 1 1  Unknown 3 1 diabetes in pregnancy 2/67 1/80 0.207 maternal smoking present (0 = nonsmoker, 1–3 = half pks/day increments) 0 = 47, 1 = 11, 2 or more = 9 0 = 60, 1 = 8, 2 or more = 12 0.721 gender of baby 27 females, 40 males 42 females, 38 males 0.352 parity = number of women 0 = 37, 1 = 20, 2 = 7, 3 or more = 3 0 = 40, 1 = 29, 2 = 7, 3 or more = 4 0.254 previous miscarriages 11/67 15/80 0.203 previous case pregnancy 1 0 0.967
Results
Participation rates Nine hundred and ninety four women were identified as eligible (342 cases and 652 controls). Of the 590 women successfully contacted, 130 (22%) agreed to participate (56 cases and 74 controls). Four hundred and four women were lost to follow-up. Cases were more likely to choose to participate than controls and this difference was significant (1.5,  p  = 0.030). There was no difference in the proportions of cases and controls that were lost to follow-up ( p  = 0.157). We had anticipated a 20% response rate and we achieved 24%.
Participation rates
Nine hundred and ninety four women were identified as eligible (342 cases and 652 controls). Of the 590 women successfully contacted, 130 (22%) agreed to participate (56 cases and 74 controls). Four hundred and four women were lost to follow-up. Cases were more likely to choose to participate than controls and this difference was significant (1.5,  p  = 0.030). There was no difference in the proportions of cases and controls that were lost to follow-up ( p  = 0.157). We had anticipated a 20% response rate and we achieved 24%.ppGenotype results Genotypes were available for 54 cases and 73 controls for this analysis. Results are summarized in Table  1 . The allele frequency for the C677T MTHFR variant in the Manitoba population has been previously established to be q = 0.25. Women who had complications of pregnancy and an unexplained MSAFP elevation had a higher allele frequency for the C677T MTHFR variant (q = 0.36) compared to women with MSAFP elevations and normal pregnancy outcomes (q = 0.25, OR 1.73 95% CI 1.25–2.37,  p  = 0.03). The frequency was also higher than in the population controls (q = 0.25, OR 1.70 95% CI 1.11–2.60,  p  = 0.007). The frequency in women without pregnancy complications and MSAFP elevations (study controls) was not significantly different than that seen in population controls ( p  = 0.41). Table 1 Comparison of allele frequency of C677T MTHFR between cases, study controls, and population controls. Subjects C/C (%) C/T (%) T/T (%) Comparing to study controls*  OR (95%CI) Comparing to population*  OR (95%CI) Cases N = 54 21 (39) 27 (50) 6 (11) 1.73 (1.25–2.37) ( p  = 0.033) 1.70 (1.11–2.60) ( p  = 0.007) Study Controls N = 73 40 (55) 30 (41) 3 (4) ~ 0.98 (0.46–1.55) ( p  = 0.410) Population N = 977 557 (57) 352 (36) 68 (7) 0.98 (0.46–1.55) ( p  = 0.410) ~ * χ 2  comparison of allele frequency (total T and C) in each group, one tailed. Cases: women with unexplained elevations of MSAFP who had subsequent complications of pregnancy, (C = 69, T = 39) Controls: women who had unexplained elevations of MSAFP and no subsequent complications (C = 110, T = 36) Population controls were 977 newborns (C = 1466, T = 488) [12]. C/C = normal type, C/T = heterozygous for thermolabile variant, T/T = homozygous for thermolabile variant. The case and study control groups included women at various stages of their child bearing years. Only one woman recruited as a control subject had a previous or subsequent pregnancy with an unexplained elevation of MSAFP and complications. She was a heterozygote for C677T MTHFR. No case subjects had a previous or subsequent pregnancy with an unexplained elevation of MSAFP and a normal outcome, but four case subjects had had a previous or subsequent pregnancy with complications after an elevated MSAFP. If the case versus control classification had been based on whether or not a woman had  ever  had a pregnancy with an unexplained elevation of MSAFP followed by complications, the association would still be present when compared to the population control (q = 0.3636, OR 1.72, 95%CI 1.27–2.61,  p  = 0.0055).
Genotype results
Genotypes were available for 54 cases and 73 controls for this analysis. Results are summarized in Table  1 . The allele frequency for the C677T MTHFR variant in the Manitoba population has been previously established to be q = 0.25. Women who had complications of pregnancy and an unexplained MSAFP elevation had a higher allele frequency for the C677T MTHFR variant (q = 0.36) compared to women with MSAFP elevations and normal pregnancy outcomes (q = 0.25, OR 1.73 95% CI 1.25–2.37,  p  = 0.03). The frequency was also higher than in the population controls (q = 0.25, OR 1.70 95% CI 1.11–2.60,  p  = 0.007). The frequency in women without pregnancy complications and MSAFP elevations (study controls) was not significantly different than that seen in population controls ( p  = 0.41).pppComparison of allele frequency of C677T MTHFR between cases, study controls, and population controls.SubjectsC/C (%)C/T (%)T/T (%)Comparing to study controls*Comparing to population*Cases N = 54ppStudy Controls N = 73pPopulation N = 977p* χ 2  comparison of allele frequency (total T and C) in each group, one tailed. Cases: women with unexplained elevations of MSAFP who had subsequent complications of pregnancy, (C = 69, T = 39) Controls: women who had unexplained elevations of MSAFP and no subsequent complications (C = 110, T = 36) Population controls were 977 newborns (C = 1466, T = 488) [12]. C/C = normal type, C/T = heterozygous for thermolabile variant, T/T = homozygous for thermolabile variant.The case and study control groups included women at various stages of their child bearing years. Only one woman recruited as a control subject had a previous or subsequent pregnancy with an unexplained elevation of MSAFP and complications. She was a heterozygote for C677T MTHFR. No case subjects had a previous or subsequent pregnancy with an unexplained elevation of MSAFP and a normal outcome, but four case subjects had had a previous or subsequent pregnancy with complications after an elevated MSAFP. If the case versus control classification had been based on whether or not a woman had  ever  had a pregnancy with an unexplained elevation of MSAFP followed by complications, the association would still be present when compared to the population control (q = 0.3636, OR 1.72, 95%CI 1.27–2.61,  p  = 0.0055).everpBiochemical results Heterozygotes and homozygotes for C677T MTHFR had lower average values (r = 0.978,  p  = 0.019) for serum folate than those who did not have the variant. None of the women were deficient in either serum folate (defined as <7.0 nmol/L) or red blood cell folate (defined as <430 nmol/L RBC). There was no significant difference in mean homocysteine levels (Table  2 ). Table 2 Comparison of the parametric characteristics of women with unexplained elevations of MSAFP according to those with and without complications of pregnancy Characteristic Mean Cases (SD) Mean Controls (SD) p  value MSAFP result 2.78 (± 0.62) 3.16 (± 3.76) 0.398 weeks gestation 17.1 (± 1.61) 16.9 (± 1.40) 0.432 μg/folate/day in pregnancy 2 1216 (± 915) 1010 (± 892) 0.206 μg/folate/day at time of study 2 557 (± 341) 523 (± 498) 0.588 erc folate (nmol/L RBC) 1234 (± 289) 1208 (± 317) 0.632 serum folate (nmol/L) 32.3 (± 5.80) 32.3 (± 5.71) 0.956 serum homocysteine (μmol/L) 7.8 (± 2.26) 8.4 (± 2.80) 0.246 μg B 12 /day in pregnancy 2 12.4 (± 5.37) 13.4 (8.31) 0.488 μg B 12 /day at time of study 2 8.9 (± 12.26) 8.6 (± 10.83) 0.899 mg B 6 /day in pregnancy 2 8.4 (± 9.69) 7.2 (± 9.01) 0.461 mg B 6 /day at time of study 2 6.0 (± 13.35) 5.5 (± 11.10) 0.792 mother's age at delivery 31 (± 4.19) 30 (± 5.19) 0.251 mother's weight in Kg 76 (± 17.26) 69 (± 16.09) 0.013 1 1  This value is not significant after Bonferroni correction for multiple comparisons. See discussion.  2 Data was skewed due to a small number of women in both groups taking large dose vitamin supplements. When these women were removed from the analysis the result remained nonsignificant.
Biochemical results
Heterozygotes and homozygotes for C677T MTHFR had lower average values (r = 0.978,  p  = 0.019) for serum folate than those who did not have the variant. None of the women were deficient in either serum folate (defined as <7.0 nmol/L) or red blood cell folate (defined as <430 nmol/L RBC). There was no significant difference in mean homocysteine levels (Table  2 ).pComparison of the parametric characteristics of women with unexplained elevations of MSAFP according to those with and without complications of pregnancyCharacteristicMean Cases (SD)Mean Controls (SD)p  valuep1  This value is not significant after Bonferroni correction for multiple comparisons. See discussion.  2 Data was skewed due to a small number of women in both groups taking large dose vitamin supplements. When these women were removed from the analysis the result remained nonsignificant.Validity of surveys Seven cases and one study control declined to fill out their dietary surveys. Mean values were inserted in the multivariate analysis for these eight women. The validity of the dietary survey was demonstrated again for this study by linear regression analysis. Consistent with known homocysteine metabolism [ 23 ], a negative correlation existed between serum homocysteine and both red blood cell folate (r = 0.945  p  = 0.0052) and serum folate (r = 0.932,  p  = 0.0001). Higher intake of dietary folate (including synthetic folic acid from supplements) as reported by the FFQ for the time of study was associated with higher serum folate (r = 0.941,  p  = 0.0001) and higher red blood cell folate (r = 0.949,  p  = 0.0166). We did several checks to determine that the women were answering their surveys accurately. Comparisons of specific data items available in the women's MMSSP charts at the time of pregnancy with the data reported in the surveys showed excellent agreement for every item examined indicating women answered questions accurately. Women who reported smoking (as a quantitative value from 0–3 based on 1/2 packs/day smoked) showed a negative correlation with serum folate (r = 0.923,  p  = 0.0062) consistent with accurately reporting their smoking habits [ 24 ]. Based on the results of these tests of the validity of our surveys, we are confident that the information provided by our participants was accurate.
Validity of surveys
Seven cases and one study control declined to fill out their dietary surveys. Mean values were inserted in the multivariate analysis for these eight women. The validity of the dietary survey was demonstrated again for this study by linear regression analysis. Consistent with known homocysteine metabolism [ 23 ], a negative correlation existed between serum homocysteine and both red blood cell folate (r = 0.945  p  = 0.0052) and serum folate (r = 0.932,  p  = 0.0001). Higher intake of dietary folate (including synthetic folic acid from supplements) as reported by the FFQ for the time of study was associated with higher serum folate (r = 0.941,  p  = 0.0001) and higher red blood cell folate (r = 0.949,  p  = 0.0166).ppppWe did several checks to determine that the women were answering their surveys accurately. Comparisons of specific data items available in the women's MMSSP charts at the time of pregnancy with the data reported in the surveys showed excellent agreement for every item examined indicating women answered questions accurately. Women who reported smoking (as a quantitative value from 0–3 based on 1/2 packs/day smoked) showed a negative correlation with serum folate (r = 0.923,  p  = 0.0062) consistent with accurately reporting their smoking habits [ 24 ]. Based on the results of these tests of the validity of our surveys, we are confident that the information provided by our participants was accurate.pAnalysis of FFQ survey and MMSSP data The ethnicity of the infants born to the case mothers (based on the ethnicity reported for the infants grandparents) was 84% Caucasian, 5% Aboriginal. Mixed ethnicity was reported for 11% of the infants with one parent Caucasian and the other parent Aboriginal, or rarely Black or Asian. The ethnic distribution was the same for controls and is typical for the Manitoba population [ 25 , 26 ]. There were also no significant differences between cases and controls with respect to their place of residence within the province (such as rural versus urban address). There were no significant differences in dietary and supplemental intake of folate, B 12 , or B 6 , or in the biochemical parameters of case and control mothers. There was no difference in the percentage of cases and controls who reported taking prenatal vitamin supplements during pregnancy (37/48 cases and 55/72) or taking vitamin and/or folate supplements preconceptionally (17/48 cases and 25/72 study controls). We attempted to divide our cases into smaller groups by type of pregnancy complication. We also separated isolated IUGR and IUGR associated with hypertensive disorders of pregnancy. Most of the groups lacked power for statistical analysis due to small numbers. However, normotensive women whose fetus had IUGR (N = 12) had a higher frequency of the C677T MTHFR variant compared to the population controls (q = 0.33, OR 2.58 95% CI, 1.78–3.73,  p  = 0.013). Homozygosity for the C677T MTHFR variant is associated with IUGR in women who do not take vitamin supplements according to one large study of Canadian women [ 10 , 27 ]. Our findings are in agreement with this result as only 3/12 women took supplements. We found this effect in a group of combined heterozygous and homozygous women. There was a trend towards higher mean weight for mothers who had complications at the time of MSAFP test in the individual comparisons, but this was not significant after correction for multiple comparisons (Table  2  and  3 ). The multivariate analysis did not reveal any unexpected associations, but it did show the importance of maternal weight as a variable (r = 0.933,  p  = 0.024). This was also not unexpected given that some of the complications we were examining are associated with obesity [ 28 ]. Even after controlling for women's weight in the multivariate analysis, the higher frequency of C677T MTHFR among cases remained significant (r = 0.734,  p  = 0.0462). There was no association between weight and MTHFR status (r = 0.431,  p  = 0.679). Table 3 Comparison of the nonparametric characteristics of women with unexplained elevations of MSAFP according to those with and without complications of pregnancy. Nonparametric Characteristics Cases Controls p  value Location  Winnipeg 44 63 0.854  southern city 3 2  southern town 3 5  southern rural 8 8  northern city 2 1  northern town 1 4  northern rural 5 7 ethnicity 0.972  Caucasian 52 66  Aborginal 2 2  Mixed Caucasian/Aboriginal 7 6  Mixed Caucasian/Black 1 3  Asian 1 1  Mixed Caucasian/Asian 1 1  Unknown 3 1 diabetes in pregnancy 2/67 1/80 0.207 maternal smoking present (0 = nonsmoker, 1–3 = half pks/day increments) 0 = 47, 1 = 11, 2 or more = 9 0 = 60, 1 = 8, 2 or more = 12 0.721 gender of baby 27 females, 40 males 42 females, 38 males 0.352 parity = number of women 0 = 37, 1 = 20, 2 = 7, 3 or more = 3 0 = 40, 1 = 29, 2 = 7, 3 or more = 4 0.254 previous miscarriages 11/67 15/80 0.203 previous case pregnancy 1 0 0.967
Analysis of FFQ survey and MMSSP data
The ethnicity of the infants born to the case mothers (based on the ethnicity reported for the infants grandparents) was 84% Caucasian, 5% Aboriginal. Mixed ethnicity was reported for 11% of the infants with one parent Caucasian and the other parent Aboriginal, or rarely Black or Asian. The ethnic distribution was the same for controls and is typical for the Manitoba population [ 25 , 26 ]. There were also no significant differences between cases and controls with respect to their place of residence within the province (such as rural versus urban address).There were no significant differences in dietary and supplemental intake of folate, B 12 , or B 6 , or in the biochemical parameters of case and control mothers. There was no difference in the percentage of cases and controls who reported taking prenatal vitamin supplements during pregnancy (37/48 cases and 55/72) or taking vitamin and/or folate supplements preconceptionally (17/48 cases and 25/72 study controls).We attempted to divide our cases into smaller groups by type of pregnancy complication. We also separated isolated IUGR and IUGR associated with hypertensive disorders of pregnancy. Most of the groups lacked power for statistical analysis due to small numbers. However, normotensive women whose fetus had IUGR (N = 12) had a higher frequency of the C677T MTHFR variant compared to the population controls (q = 0.33, OR 2.58 95% CI, 1.78–3.73,  p  = 0.013). Homozygosity for the C677T MTHFR variant is associated with IUGR in women who do not take vitamin supplements according to one large study of Canadian women [ 10 , 27 ]. Our findings are in agreement with this result as only 3/12 women took supplements. We found this effect in a group of combined heterozygous and homozygous women.pThere was a trend towards higher mean weight for mothers who had complications at the time of MSAFP test in the individual comparisons, but this was not significant after correction for multiple comparisons (Table  2  and  3 ). The multivariate analysis did not reveal any unexpected associations, but it did show the importance of maternal weight as a variable (r = 0.933,  p  = 0.024). This was also not unexpected given that some of the complications we were examining are associated with obesity [ 28 ]. Even after controlling for women's weight in the multivariate analysis, the higher frequency of C677T MTHFR among cases remained significant (r = 0.734,  p  = 0.0462). There was no association between weight and MTHFR status (r = 0.431,  p  = 0.679).pppComparison of the nonparametric characteristics of women with unexplained elevations of MSAFP according to those with and without complications of pregnancy.Nonparametric CharacteristicsCasesControlsp  valuepDiscussion Unexplained elevations in MSAFP are known to be associated with an increased risk for complications of pregnancy [ 2 ]. Others have reported that presence of the C677T MTHFR variant in pregnant women with low folate intake is associated with increased risk for pregnancy complications [ 2 , 29 - 31 ]. The unique finding of this study is an increase in the frequency of the C677T MTHFR variant among women with normal folate intake, who went on to have complications of pregnancy after an unexplained elevation of MSAFP (Table  1 ). The lack of folate deficiency in this population was unexpected, given previous research which showed that 23.6% of Newfoundland and Labrador women are folate deficient at their first prenatal visit [ 32 ]. As our study was retrospective, we did not have data on levels during pregnancy. It has recently been shown that the C677T MTHFR variant does not affect maternal serum homocysteine levels in pregnancy among women who take prenatal multivitamins [ 8 ]. Also a recent prospective study shows that there is no difference in homocysteine levels at midtrimester between women who later develop preeclampsia and those who do not [ 33 ]. As is the situation with NTDs, lack of folate deficiency by current definitions in a non-pregnant woman may not indicate that her folate intake is adequate for pregnancy. This would especially be true for women with the C677T MTHFR variant. Reexamination of the current definition of what constitutes a normal biochemical result for folate intake for women of child bearing age should be undertaken to clarify this. We suggest that the negative effects of the C677T MTHFR variant are more likely to occur in early pregnancy before women began taking prenatal vitamins because the majority of our study participants took prenatal vitamins, but only 35% took preconceptional supplements. We suspect that reduced methylation interfering with cell proliferation in the placenta as originally suggested by Eskes (2000) [ 3 ]. In conclusion, using a retrospective case/control study, we have found that women with unexplained MSAFP elevations who have complications in later pregnancy are more likely to have the C677T MTHFR allele. Our resultsdo not suggest that C677T MTHFR predisposes a woman to having an elevation of MSAFP level (as we did not compare the C677T MTHFR frequency in women with and without elevated MSAFP), but having one or more copies of this variant predisposes such screen positive women to having complications in later gestation. It remains to be seen if other risk factors can be identified which can more accurately define this high risk group.
Discussion
Unexplained elevations in MSAFP are known to be associated with an increased risk for complications of pregnancy [ 2 ]. Others have reported that presence of the C677T MTHFR variant in pregnant women with low folate intake is associated with increased risk for pregnancy complications [ 2 , 29 - 31 ]. The unique finding of this study is an increase in the frequency of the C677T MTHFR variant among women with normal folate intake, who went on to have complications of pregnancy after an unexplained elevation of MSAFP (Table  1 ).The lack of folate deficiency in this population was unexpected, given previous research which showed that 23.6% of Newfoundland and Labrador women are folate deficient at their first prenatal visit [ 32 ]. As our study was retrospective, we did not have data on levels during pregnancy. It has recently been shown that the C677T MTHFR variant does not affect maternal serum homocysteine levels in pregnancy among women who take prenatal multivitamins [ 8 ]. Also a recent prospective study shows that there is no difference in homocysteine levels at midtrimester between women who later develop preeclampsia and those who do not [ 33 ].As is the situation with NTDs, lack of folate deficiency by current definitions in a non-pregnant woman may not indicate that her folate intake is adequate for pregnancy. This would especially be true for women with the C677T MTHFR variant. Reexamination of the current definition of what constitutes a normal biochemical result for folate intake for women of child bearing age should be undertaken to clarify this.We suggest that the negative effects of the C677T MTHFR variant are more likely to occur in early pregnancy before women began taking prenatal vitamins because the majority of our study participants took prenatal vitamins, but only 35% took preconceptional supplements. We suspect that reduced methylation interfering with cell proliferation in the placenta as originally suggested by Eskes (2000) [ 3 ].In conclusion, using a retrospective case/control study, we have found that women with unexplained MSAFP elevations who have complications in later pregnancy are more likely to have the C677T MTHFR allele. Our resultsdo not suggest that C677T MTHFR predisposes a woman to having an elevation of MSAFP level (as we did not compare the C677T MTHFR frequency in women with and without elevated MSAFP), but having one or more copies of this variant predisposes such screen positive women to having complications in later gestation. It remains to be seen if other risk factors can be identified which can more accurately define this high risk group.Authors' contributions All authors participated in original study design except CS. CG, BC and CS all acted as principal investigators for funding. NB assisted with all grant proposals. NB undertook the review of the individual MSAFP files and MMSSP database searches, designed the dietary and family history surveys, classified cases and controls, acted as study coordinator handling all aspects of participant contact, and provided data analysis. All authors also participated actively with NB for various aspects of the study. CG provided MTHFR genotyping. LS provided biochemical analysis. NB drafted the original manuscript with assistance from BC. CG and BC handled ethics approval assisted by NB. All authors read and approved the final manuscript.
Authors' contributions
All authors participated in original study design except CS. CG, BC and CS all acted as principal investigators for funding. NB assisted with all grant proposals. NB undertook the review of the individual MSAFP files and MMSSP database searches, designed the dietary and family history surveys, classified cases and controls, acted as study coordinator handling all aspects of participant contact, and provided data analysis. All authors also participated actively with NB for various aspects of the study. CG provided MTHFR genotyping. LS provided biochemical analysis. NB drafted the original manuscript with assistance from BC. CG and BC handled ethics approval assisted by NB. All authors read and approved the final manuscript.