Exclusive breastfeeding without supplementation is recommended for the first 6 months of life by many authorities, including the World Health Organization1 and the American College of Obstetricians and Gynecologists.2–5 Breast milk confers immunity, is nutritionally balanced, easily digested, and promotes healthy growth.1 Breastfeeding is associated with lower rates of disease not only in newborns (sudden infant death syndrome, respiratory, gastrointestinal, and ear infections) and children (allergies, asthma, obesity), but also mothers (breast cancer and ovarian cancer).6
Supplementation in the early postpartum period is associated with early cessation of breastfeeding.7–10 Supplementation before hospital discharge is associated with shorter breastfeeding duration (adjusted odds ratio [OR] 3.9, 95% confidence interval [CI] 2.1–7.29), and lack of supplementation is associated with longer breastfeeding duration (OR 2.49, 95% CI 1.25–4.9810). Exclusive breastfeeding at hospital discharge is associated with a markedly lower risk of weaning by 6 months postpartum (adjusted OR 0.19, 95% CI 0.06–0.65, P=.008).11
Despite the recognized benefits of breastfeeding, current knowledge about the prevalence and predictors of breastfeeding is inadequate as a result of major methodologic limitations of existing studies, which include: volunteer bias, selection bias, recall bias, and lack of reporting of gestational ages.12 Additional, methodologically sound research was called for to determine an unbiased rate of breastfeeding.12
Understanding breastfeeding requires an exploration beyond solely maternal and newborn factors that could affect exclusive breastfeeding; delivery, health care provider, and hospital factors also might affect breastfeeding. Our objectives were to estimate population-based rates of exclusive breastfeeding at maternal–newborn discharge from the hospital and to understand maternal, antenatal, health care provider, hospital, delivery, and newborn factors associated with exclusive breastfeeding in term newborns in a population-based study.
MATERIALS AND METHODS
We performed a retrospective population-based cohort study following the guidelines for the conduct of cohort studies (Strengthening the Reporting of Observational Studies in Epidemiology).13 Between April 1, 2009, and March 31, 2010, we identified all births in all Ontario hospitals.
We used a population-based database that captures information on all hospital births in the province of Ontario, Canada, the most populous province in the country.14 It is termed the Better Outcomes Registry & Network, of which the Niday Perinatal Database is the hub.15 The Better Outcomes Registry & Network is a Prescribed Registry under Ontario's Personal Health Information Protection Act. High data quality is maintained by formal training of new users and logic-checking mechanisms that are built into the system to minimize missing or erroneous data. Each hospital receives quality reports and the data are edited to correct errors. A recent quality audit of the Better Outcomes Registry & Network data found that the percent agreement compared with chart abstraction for variables we included was typically in the mid- to high 90s except for episiotomy at 82.7%, forceps or vacuum at 86.5%, laceration at 75%, smoking at 78.9%, and nonpharmacologic pain methods at 77.1%.16 Antepartum information is entered in prebirth registration clinics and on labor floors before birth, whereas postpartum information such as feeding at discharge is entered by nurses on the postpartum floors before maternal and newborn discharge.
We included live singleton and twin births at term (37 0/7 weeks of gestation to 41 6/7 weeks of gestation) with information about feeding at maternal–newborn discharge. We excluded all triplets and higher-order pregnancies, stillborn neonates, neonatal deaths, newborns with congenital anomalies, newborns discharged to a location other than home (another hospital) or from a neonatal intensive care unit or special care unit (because their information is not within the database), and planned home births, which are not included in Niday. We randomly excluded one twin as a result of a high correlation of outcomes within twins.
Our primary outcome was exclusive breastfeeding (without formula supplementation) at maternal–newborn discharge from the hospital. The potential predictor variables were grouped into maternal, antenatal, health care provider, hospital, delivery, and newborn factors. We included the following maternal and antenatal predictor variables: age, number of previous term and preterm neonates, any smoking during pregnancy or any illicit drug use, physical or mental health problems, twin pregnancy, use of assisted reproductive technology, obstetric or pregnancy complication (one or more of the following during the antepartum: gestational diabetes, hypertension, suspected intrauterine growth restriction or small for gestational age, suspected large for gestational age, periodontal infection, placental abruption, preeclampsia, premature rupture of membranes, preterm labor, preterm premature rupture of membranes, urinary tract infection), attendance at prenatal classes, and type of antenatal care provider (exclusive care by: midwife, family physician, nurse practitioner, or obstetrician or no exclusive care provider). Although individual participant income and employment were not available, we did have the participants' median neighborhood income entered in $20,000 increments in the regression model and the median proportion employed in their neighborhood and in the labor force. (These data were defined according to Statistics Canada census data17 based on the postal code, a six-character code, defined and maintained by the national postal system for the purpose of sorting and delivering mail.) The type of hospital was characterized by the level of care designations,18 which are defined based on the most acute level of care a hospital would typically provide: Level 1: healthy, singletons 36 weeks of gestation or greater, 2,500 g or greater, or both; Level 2: mothers and newborns at low-to-moderate risk 32 weeks of gestation or greater; Level 2+: mothers and newborns at moderate-risk, singletons 28–30 weeks of gestation or greater, or twins 34 weeks of gestation or water; and Level 3: mothers with severe medical complications, unstable newborns, high-risk pregnancies including triplets and higher-order multiples, and newborns of any gestational age. Delivery predictor variables included: maternal pain relief (none, local anesthetic, pudendal nerve block, narcotics, nitrous oxide, nonpharmacologic methods, spinal epidural, or general anesthetic), intrapartum complications (bleeding, meconium, nonprogressive labor, suspected chorioamnionis, suspected sepsis, cord prolapse, uterine rupture, shoulder dystocia, postpartum hemorrhage), and type of delivery (spontaneous vaginal delivery, vaginal delivery [induction or augmentation], assisted vaginal delivery [forceps, vacuum, or episiotomy], planned cesarean delivery and unplanned cesarean delivery). Newborn predictor variables included number of weeks of gestation at delivery, birth weight, newborn sex, Apgar score at 5 minutes, umbilical arterial cord pH, and newborn resuscitation (more than one of: free flow oxygen, positive pressure ventilation, intubation, chest compressions, resuscitative medications).
Baseline descriptive statistics were performed to characterize women who exclusively breastfed and those who did not. Each potential predictor variable (maternal, antenatal, health care provider, hospital and delivery, and newborn variables) was first examined in a univariable fashion using logistic regression. Then in a multivariable logistic regression model, we adjusted for maternal and antenatal, hospital and delivery, and newborn variables to yield adjusted ORs for exclusive breastfeeding. Because breastfeeding outcomes were almost identical within twin pairs, we randomly selected one newborn from each pair of twins to examine. A single fiscal year of data was used to minimize the possibility of any given woman having more than one pregnancy included in the data. Missing predictor variables were imputed using multiple imputation.19,20 As a sensitivity analysis, we compared results with and without multiple imputation and found no important differences.
A priori we performed a sample size calculation. We conservatively estimated that approximately 100,000 term births per year occur in Ontario. With 95% of births occurring in the hospital, and with an estimate that data on feeding at discharge would be available for approximately 80% of births, we expected to have 100,000×0.95×0.80=76,000 newborns available for the analyses. We estimated that approximately 40% of all singletons and twins would be exclusively breastfed at discharge. A two-tailed α of 5% was used in power calculations. Power to detect predictors of exclusive breastfeeding depends on the relative sizes of the groups to be compared, which differs across predictors, but we estimated that, for example, with twins making up 1.4% of births, we would have 80% power to detect a relative risk of 0.77 for twins compared with singletons. Thus, even with a relatively rare occurrence such as twins, we felt confident that even 1 year's worth of data would be statistically sufficiently large to detect important differences. Moreover, allowing for 10 breastfeeding “events” per variable in the regression models, for the primary analysis, this would be 76,000×0.4/10=3,040, which is many more covariates than would ever be usefully included in a model.
Analyses were performed using SAS-PC 9.2 statistical software. The study was approved by the Faculty of Health Sciences, McMaster University research ethics board.
Of the 140,322 neonates born between April 1, 2009, and March 31, 2010, in hospitals in Ontario, Canada (Fig. 1), there were 114,680 singletons or twin newborns born at term without congenital anomalies. Of these, there were 22,316 (15.9% of 143,322) newborns missing information on feeding at discharge, yielding 92,364 newborns who constituted our study population. Baseline characteristics of the women with and without information on feeding at discharge were similar on all compared characteristics (maternal ages 29.8 compared with 30.6 years, respectively, neighborhood median family incomes $72,794 compared with $73,650, mean number of previous pregnancies 0.89 compared with 0.89, and median gestational ages at delivery 39 compared with 39 weeks of gestation). Baseline descriptive characteristics of the 92,364 newborns who constituted our study population are presented in Table 1, of whom 56,865 (61.6%) newborns were exclusively breastfed at discharge and 35,499 (38.5%) were not. Maternal, antenatal, health care provider, hospital, delivery, and newborn factors were compared between the two groups.
The adjusted ORs for initiation of exclusive breastfeeding are shown in Table 2 with important ones highlighted here. Older women had higher odds of exclusive breastfeeding (OR 1.26, 95% CI 1.23–1.30 for 10-year increments) as did women who lived in areas that have higher family incomes (OR 1.08 95% CI 1.07–1.10 for $20,000 increments). Having had a previous term (OR 0.96, CI 0.95–0.98) or preterm (OR 0.83, 95% CI 0.79–0.87) birth was associated with slight decreases in exclusive breastfeeding. Women who did not have drug problems (OR 1.41, 95% CI 1.31–1.52), smoke (OR 1.63, 95% CI 1.56–1.70), have obstetric or pregnancy complications (OR 1.21, 95% CI 1.17–1.25), or reproductive assistance (OR 1.28, 95% CI 1.14–1:43) were more likely to exclusively breastfeed their newborns. Mothers of twins (OR 0.30, 95% CI 0.25–0.36) and women who did not attend prenatal classes (OR 0.80, 95% CI 0.76–0.83) were less likely to exclusively breastfeed.
Compared with women whose antenatal care was provided exclusively by obstetricians, women receiving exclusive midwifery care had 4.5 times the odds of exclusively breastfeeding (87% compared with 57%, OR 4.49, 95% CI 4.16–4.85). Women cared for by family physicians (67%, OR 1.54, 95% CI 1.47–1.61) and those with no exclusive care by any one type of care provider (63%, OR 1.33, 95% CI 1.29–1.38) were also more likely to exclusively breastfeed compared with women cared for by obstetricians.
Compared with women in level III hospitals, women delivering in level II hospitals (OR 1.11, 95% CI 1.05–1.17) were more likely to exclusively breastfeed, whereas those delivering in level II+ (OR 0.83, 95% CI 0.79–0.88) were less likely. Women with either a planned cesarean delivery (50%, OR 0.56, 95% CI 0.52–0.60) or an unplanned cesarean delivery (48%, OR 0.48, 95% CI 0.44–0.51) were less likely than women with spontaneous vaginal deliveries to exclusively breastfeed (68%). Women without intrapartum complications (OR 1.10, 95% CI 1.06–1.14) and with no or minimal pain relief (OR 1.17, 95% CI 1.13–1.21) were more likely to exclusively breastfeed.
Although we only considered newborns delivered at term, women delivering at 39, 38, and 37 weeks of gestation were increasingly less likely to exclusively breastfeed compared with women delivering at 41 weeks of gestation (OR 0.93, 95% CI 0.89–0.98; OR 0.84, 95% CI 0.80–0.88; and OR 0.71, 95% CI 0.67–0.76, respectively). Neonates with higher Apgar scores (1.07, 95% CI 1.04–1.10) and who did not need resuscitation were slightly more likely (OR 1.07, 95% CI 1.03–1.12) to be exclusively breastfed.
We found that, despite recommendations, fewer than two thirds of term newborns are exclusively breastfed when discharged from the hospital with their mothers. Given both the steep decline in breastfeeding that occurs in the first few months postpartum,21,22 and the markedly higher risk of weaning by 6 months when newborns are not exclusively breastfed at hospital discharge,11 the comparatively low proportion of women who exclusively breastfeed at discharge from the hospital is very concerning.
There is very little true population-based information on breastfeeding with the exception of data from 20 years ago, from Avon, United Kingdom, which reported a higher rate (76%) of exclusive breastfeeding at discharge from the hospital.21 Most data are based on women who agreed to participate in a survey and thus are vulnerable to volunteer bias and recall bias,23 including surveys from Poland, which found 69% of newborns were exclusively breastfed at discharge from the hospital in 1995,22 Italy (91% in 1999),24 Australia (82% in 1995),25 France (43–80% depending on the region in 2003),26 and Canada (85% in 2005).27 Given that surveys suggest marked variation in self-reported breastfeeding from country to country, the prevalence of breastfeeding according to population-based data may also vary from country to country.
Our large sample size allowed exploration of novel variables such as the week of gestational age at delivery, demonstrating profound effects within a time period that is considered to be term in clinical practice. Newborns born at 37 weeks of gestation had only 71% of the odds of being exclusively breastfed compared with newborns born at 41 weeks of gestation even after controlling for other factors. Although previous studies have documented that cesarean delivery is associated with lower rates of breastfeeding,28 we were able to subdivide cesarean delivery and found exclusive breastfeeding even less likely after unplanned cesarean delivery than planned. Although some women or newborns who experience cesarean delivery may be unwell, often, hospital practices may be less likely to support skin-to-skin contact and attempts to latch on within the first hour after birth after cesarean delivery.
Our findings add to the literature examining nonmaternal and newborns factors influencing exclusive breastfeeding, including type of care provider and hospital. Although part of the profoundly higher odds of exclusive breastfeeding seen in women cared for by midwives is the result of self-selection of the type of woman who seeks midwifery care (a relatively new model of care since 1994 in Ontario), part may be attributed to the care providers' support of breastfeeding. Women who received antenatal care from family physicians and nurse practitioners also demonstrated higher odds of breastfeeding compared with those of obstetricians. Previous studies reported that 64% of nurse midwives in Iowa, 13% of family practitioners, and 7% of obstetricians, self-reported that they were strong breastfeeding advocates.29 Although one study suggested that advice and support from the hospital were not associated with maintenance of breastfeeding,30 we found higher exclusive breastfeeding in level II hospitals (designed to care for neonates born at 32 weeks of gestation or greater) than level III hospitals. Further exploration of why these differences exist within varying levels of care would potentially allow systems' modifications to better support breastfeeding. Recent work has suggested relatively low levels of compliance with the World Health Organization's and UNICEF's Baby Friendly Initiative,31 with more than one third of Canadian women reporting being offered samples of formula and almost half of women reporting that they offered their newborn a pacifier or soother within the first week of life.32 Audits of the 10 steps to promote breastfeeding would yield important information on systems' issues for targeted intervention.
Other strengths of our study include the population-based design with data from Ontario, Canada's most populous province. In addition, the database's strengths include real-time data entry, information on a broad range of maternal and newborn factors as well as data on health care providers and hospitals. Because of the large sample size, we were able to examine subcategories of factors such as week at delivery and rare factors such as twins.
Our study has limitations including the fact that we did not have information on some factors such as family support or ethnicity, which might influence breastfeeding. Some variables had incomplete information, for instance, body mass index, and hence were not included in the analysis. Although approximately 16% of women did not have information about feeding at discharge, baseline characteristics were similar between women with these data to those with missing data. The Niday database does not contain information after discharge.
In summary, these current, population-based data reveal that exclusive breastfeeding rates are suboptimal despite the recognized benefits for newborns, children, and mothers. Identifying predictors of breastfeeding was the critical first step in developing breastfeeding support strategies to address maternal and newborn, health care provider, and hospital barriers to improve breastfeeding. Translating the findings to health care providers and administrators will be an important next step. Future research will be required to identify strategies at each specific level to improve breastfeeding rates.
1. World Health Organization [WHO]. Report of the expert consultation on the optimal duration of exclusive breastfeeding. 182137. Geneva (Switzerland): WHO; 2001.
2. Breastfeeding: maternal and infant aspects. ACOG Committee Opinion No. 361. American College of Obstetricians and Gynecologists. Obstet Gynecol 2007;109:479–80.
3. Gartner LM, Morton J, Lawrence RA, Naylor AJ, O'Hare D, Schanler RJ, et al.. Breastfeeding and the use of human milk. Pediatrics 2005;115:496–506.
4. Boland M. Exclusive breastfeeding should continue to six months. Paediatr Child Health 2005;10:148.
5. Royal College of Paediatrics and Child Health. Position statement—breastfeeding. Available at: www.rcpch.ac.uk/child-health/standards-care/position-statements/position-statements
. 2001. Retrieved October 30, 2011.
6. Ip S, Chung M, Raman G, Chew P, Magula N, DeVine D, et al.. Breastfeeding and maternal and infant health outcomes in developed countries. Evid Rep Technol Assess (Full Rep) 2007;153:1–186.
7. Holmes AV, Auinger P, Howard CR. Combination feeding of breast milk and formula: evidence for shorter breast-feeding duration from the National Health and Nutrition Examination Survey. J Pediatr 2011;159:186–91.
8. Riva E, Banderali G, Agostoni C, Silano M, Radaelli G, Giovannini M. Factors associated with initiation and duration of breastfeeding in Italy. Acta Paediatr 1999;88:411–5.
9. Blomquist HK, Jonsbo F, Serenius F, Persson LA. Supplementary feeding in the maternity ward shortens the duration of breast feeding. Acta Paediatr 1994;83:1122–6.
10. Sheehan D, Bridle B, Hillier T, Feightner K, Hayward S, Lee KS, et al.. Breastfeeding outcomes of women following uncomplicated birth in Hamilton-Wentworth. Can J Public Health 1999;90:408–11.
11. Frota DA, Marcopito LF. Breastfeeding among teenage and adult mothers in Brazil [in Portuguese]. Rev Saude Publica 2004;38:85–92.
12. Callen J, Pinelli J. Incidence and duration of breastfeeding for term infants in Canada, United States, Europe, and Australia: a literature review. Birth 2004;31:285–92.
13. von Elm E, Altman DG, Egger M, Pocock SJ, Gøtzsche PC, Vandenbroucke JP; STROBE Initiative. The Strengthening the Reporting of Observational Studies in Epidemiology (STROBE) statement: guidelines for reporting observational studies. Lancet 2007;370:1453–7.
14. Perinatal Partnership Program of Eastern and Southeastern Ontario (PPPESO). Annual perinatal statistical report 2006–07. Available at: www.nidaydatabase.com/info/pdf/200607AnnualReportFinal.pdf
. Retrieved July 11, 2011.
15. The Niday Perinatal Database. Better Outcomes Registry & Network (BORN) Ontario. Available at: www.nidaydatabase.com
. 2011. Retrieved October 30, 2011.
16. Dunn S, Bottomley J, Ali A, Walker M. 2008 Niday Perinatal Database quality audit: report of a quality assurance project. Chronic Dis Inj Can 2011;32:32–42.
17. Postal Code Conversion File (PCCF) 2006 Reference guide. Statistics Canada. Available at: www.statcan.gc.ca/pub/92f0153g/92f0153g2007001-eng.pdf
. , 2007. Retrieved July 5, 2011.
18. Niday Perinatal Database for the GTA: Fourth Annual Statistical Report 2006–2007, A report by the Child Health Network for the Greater Toronto Area, October 2007. Available at: www.nidaydatabase.com/info/pdf/Niday_Annual_Report_Final.pdf
. Retrieved July 11, 2011.
19. Rubin DB. Multiple imputation for nonresponse in surveys. New York (NY): Wiley; 1987.
20. Little RJA, Rubin DB. Statistical analysis with missing data. 2nd ed. New York (NY): Wiley; 2002.
21. Donath SM, Amir LH. Relationship between prenatal infant feeding intention and initiation and duration of breastfeeding: a cohort study. Acta Paediatr 2003;92:352–6.
22. Mikiel-Kostyra K, Mazur J, Wojdan-Godek E. Factors affecting exclusive breastfeeding in Poland: cross-sectional survey of population-based samples. Soz Praventivmed 2005;50:52–9.
23. Yang Q, Wen SW, Dubois L, Chen Y, Walker MC, Krewski D. Determinants of breast-feeding and weaning in Alberta, Canada. J Obstet Gynaecol Can 2004;26:975–81.
24. Giovannini M, Banderali G, Radaelli G, Carmine V, Riva E, Agostoni C. Monitoring breastfeeding rates in Italy: national surveys 1995 and 1999. Acta Paediatr 2003;92:357–63.
25. Donath S, Amir LH. Rates of breastfeeding in Australia by state and socioeconomic status: evidence from the 1995 National Health Survey. Breastfeed Rev 2000;8:23–7.
26. Bonet M, Blondel B, Khoshnood B. Evaluating regional differences in breast-feeding in French maternity units: a multi-level approach. Public Health Nutr 2010;13:1946–54.
27. Millar WJ, Maclean H. Breastfeeding practices, health reports. Vol 16. Catalogue no. 82-003XIE. Ottawa (Ontario, Canada): Statistics Canada; 2005. p. 23–31.
28. Scott JA, Binns CW, Graham KI, Oddy WH. Temporal changes in the determinants of breastfeeding initiation. Birth 2006;33:37–45.
29. Dusdieker LB, Dungy CI, Losch ME. Prenatal office practices regarding infant feeding choices. Clin Pediatr (Phila) 2006;45:841–5.
30. Bruce NG, Khan Z, Olsen ND. Hospital and other influences on the uptake and maintenance of breast feeding: the development of infant feeding policy in a district. Public Health 1991;105:357–68.
31. World Health Organization (WHO)/UNICEF. Ten steps to successful breastfeeding. Available at: www.unicef.org/newsline/tensps.htm
. Retrieved July 5, 2011.
32. Chalmers B, Levitt C, Heaman M, O'Brien B, Sauve R, Kaczorowski J Breastfeeding rates and hospital breastfeeding practices in Canada: a national survey of women. Birth 2009;36:122–32.