Abstract
Objectives. Commercial hospital discharge packs are commonly given to new mothers at the time of newborn hospital discharge. We evaluated the relationship between exclusive breastfeeding and the receipt of commercial hospital discharge packs in a population-based sample of Oregon women who initiated breastfeeding before newborn hospital discharge.
Methods. We analyzed data from the 2000 and 2001 Oregon Pregnancy Risk Assessment Monitoring System (PRAMS), a population-based survey of postpartum women (n=3895; unweighted response rate=71.6%).
Results. Among women who had initiated breastfeeding, 66.8% reported having received commercial hospital discharge packs. We found that women who received these packs were more likely to exclusively breastfeed for fewer than 10 weeks than were women who had not received the packs (multivariate adjusted odds ratio=1.39; 95% confidence interval=1.05, 1.84).
Conclusions. Commercial hospital discharge packs are one of several factors that influence breastfeeding duration and exclusivity. The distribution of these packs to new mothers at hospitals is part of a longstanding marketing campaign by infant formula manufacturers and implies hospital and staff endorsement of infant formula. Commercial hospital discharge pack distribution should be reconsidered in light of its negative impact on exclusive breastfeeding.
The American Academy of Pediatrics recommends exclusive breastfeeding until an infant is 6 months old.1 Numerous studies have demonstrated the beneficial effects of breastfeeding, including decreased risk of infectious diseases (e.g., diarrhea, ear infections, and respiratory infections) and chronic diseases (e.g., asthma, allergies, and obesity).2–10 Despite the well-documented evidence that supports breastfeeding, only 66% of US women initiate breastfeeding and only 33% exclusively or partially breast-feed for 6 months.11 These figures fall short of the Healthy People 2010 goals to increase the proportion of women who initiate breast-feeding to 75% and who breastfeed for at least 6 months to 50%.12
Since the late 19th century, infant formula manufacturers have encouraged mothers to substitute formula for breastmilk. Formula advertisements often claimed or implied that breastmilk alone was not sufficient to raise a healthy infant.13–17 For more than 40 years, formula manufacturers have supplied US hospitals with free formula and newborn starter pack gifts (most of which contain either formula or coupons for formula) for distribution to new mothers.17–18 These free starter packs are an efficient and effective marketing method by which formula manufacturers get new mothers to try their company’s formula.
Formula manufacturers also have sought to create partnerships and brand loyalty with hospitals and their staff by providing free formula for use in the hospital, support for fellowships and conferences, and funds to support supplies.17–20 These “gifts” have strings attached, as noted by the ethics committee of one hospital that blocked the routine distribution of free formula company discharge packs because the members viewed such distribution as distorting informed consent, prioritizing financial issues above patient care, exploiting some women’s fear of inadequacy, and implying medical endorsement of formula.21
The 1970s boycott of Nestlé (because of the company’s aggressive marketing of formula, especially in developing countries) led to international discussions about the role of formula manufacturers and ways in which hospitals could increase support for breast-feeding. These discussions culminated in the Baby-Friendly Hospital Initiative, a 1991 codification of practices by the World Health Organization. Some of the Initiative’s methods have been incorporated into routine practice in US hospital nurseries. In a 1998 study of newborn hospital breastfeeding support practices in Oregon, we found that more than 60% of Oregon newborn hospital nurseries reported moderate or high compliance with some Baby-Friendly hospital practices, such as providing rooming-in (baby stays in mother’s hospital room rather than nursery) on a routine basis, encouraging breastfeeding on demand, and refraining from offering pacifiers to newborns. Hospital practice compliance was low, however, for supplementation (including providing mothers with formula promotion items as well as giving infants formula or water).22 New mothers who responded to the 1998–1999 Oregon Pregnancy Risk Assessment Monitoring System (PRAMS) survey reported that rooming-in and breastfeeding on demand were common practices (94% and 84%, respectively) in Oregon hospitals and birthing centers. However, only 27% of the women who responded to the PRAMS survey reported that they had not received a commercial hospital discharge pack (CHDP) that contained formula.23
We sought to estimate the proportion of new mothers in Oregon who received CHDPs after initiation of breastfeeding and to examine the association between receipt of CHDPs and exclusive breastfeeding duration.
We based our study on data from the Oregon PRAMS survey, an ongoing, population-based survey of postpartum women conducted by the state public health department. The protocols for these surveys were modeled after a multistate survey supported by the Centers for Disease Control and Prevention, but data from the 2000 and 2001 Oregon PRAMS surveys were not collected under a Centers for Disease Control and Prevention protocol. Briefly, Oregon PRAMS is a cross-sectional population-based survey of a stratified systematic sample of Oregon-resident mothers who delivered a live-born infant in Oregon. Birth certificates were the source for the sampling frame, with an over-sampling of racial/ethnic minorities and non-Hispanic White women who gave birth to a low-birthweight infant. The survey employed a mixed-mode response (mail and telephone). Responses were weighted for oversampling, nonresponse, and noncoverage to be representative of the state’s entire population of women who delivered live-born infants. Details of the Oregon PRAMS methods appear elsewhere.24
We analyzed Oregon PRAMS data for the years 2000 and 2001 (inclusive of infants born from January 1, 2000, through November 4, 2001). The median number of days from birth to survey response was 104. Of the 5440 women selected to participate in the survey, 3895 completed the survey for an unweighted response rate of 71.6% (weighted response rate = 78.8%). Receipt of a CHDP was the only factor about the marketing of formula contained on the PRAMS survey. Therefore, we determined whether women who responded to the survey received CHDPs from their responses to the statement: “The staff [at the hospital or birthing center where your new baby was born] gave you a gift pack with formula.” Breastfeeding initiation was determined by responses to the question: “Did you ever breastfeed or pump breastmilk to feed your new baby after delivery?” Duration of exclusive breastfeeding was determined by responses to the question: “How old was your baby the first time you fed him or her anything besides breastmilk? Include formula, baby food, juice, cow’s milk, water, sugar water, or anything else.”
In addition to the primary independent variable of interest (receipt of a CHDP), we also analyzed demographic and prenatal characteristics of respondents that might have been associated with exclusive breastfeeding. The PRAMS datasets were the source for duration and exclusivity of breastfeeding, annual prepregnancy family income, maternal prepregnancy body mass index (weight in kilograms divided by height in meters squared), and smoking status at the time of the survey. Birth certificate data were used to obtain maternal age, education, and race/ethnicity; parity; marital status; and enrollment in the Special Supplemental Nutrition Program for Women, Infants, and Children (WIC) program during pregnancy.
Of the 3895 respondents, 1211 were excluded from analysis for the following reasons: infant was deceased or no longer with the birth mother (n = 68; per protocol, these respondents were not asked questions about breastfeeding); breastfeeding was not initiated (n = 363); failed to respond to the PRAMS question concerning the length of time the infant was breastfed (n = 657); and failed to respond to the PRAMS question about receipt of a CHDP (n = 123). The final sample for analysis included responses from 2684 women.
We analyzed responses using SPSS version 14.0 (SPSS Inc, Chicago, Ill), SUDAAN version 9.0 (Research Triangle Institute, Research Triangle Park, NC), and SAS version 9.1.3 (SAS Institute Inc, Cary, NC). We used SUDAAN and SAS-callable SUDAAN to account for the complex sample design, which involved a stratified, weighted sample. We examined distributions and frequencies of variables in SPSS. We examined variables singly for an association with exclusive breastfeeding for at least 10 weeks by the use of 2 × 2 cross-tabulations to determine odds ratios and their corresponding 95% confidence intervals. We examined potential confounders and effect modifiers through single-factor stratified analyses. We used logistic regression to study the relationship between sustained exclusive breastfeeding for at least 10 weeks and the receipt of CHDPs. We performed a similar logistic regression analysis to study the relationship between the receipt of CHDPs and nonexclusive breastfeeding for at least 10 weeks. We estimated variable significance, using weighted data, with the Wald F test statistic, setting the level of significance at less than .05. We used forward stepwise regression procedures to identify the variables with the greatest influence on sustained exclusive breastfeeding and nonexclusive breastfeeding. The final multiple variable models retained the primary independent variable of interest (receipt of a CHDP) as well as all other independent variables that were statistically significant.
Among PRAMS respondents who initiated breastfeeding, 66.8% reported having received a CHDP from the hospital. When we explored the effect of receipt of a CHDP on duration of exclusive breastfeeding among women who initiated breastfeeding, we found that women who received a CHDP were more likely to exclusively breastfeed for shorter durations than were women who did not receive a CHDP (2 weeks postpartum: odds ratio [OR] = 1.51, 95% confidence interval [CI] = 1.11, 2.05; 6 weeks postpartum: OR = 1.41, 95% CI = 1.08, 1.85; 10 weeks postpartum: OR = 1.40, 95% CI = 1.08, 1.83. [Data not shown.]). Table 1 shows the distribution of selected characteristics of the survey respondents according to whether they had exclusively breastfed their infants for at least 10 weeks duration. We found that women who exclusively breastfed for at least 10 weeks differed on the basis of maternal age, race/ethnicity, education, family income level, marital status, tobacco use, parity, and receipt of a CHDP. No differences were found between women who exclusively breastfed for 10 or more weeks and women who breastfed for less than 10 weeks based on participation in WIC or on maternal prepregnancy body mass index.
Table 2 shows the bivariate and multivariate relative odds of exclusive breastfeeding for less than 10 weeks associated with selected maternal characteristics. After we adjusted for maternal age, race/ethnicity, education, and family income, we found that women who received a CHDP were more likely to exclusively breastfeed their infants for less than 10 weeks than were women who did not receive a CHDP (adjusted odds ratio [AOR] = 1.39; 95% CI = 1.05, 1.84).
We also evaluated the effect of CHDPs on nonexclusive breastfeeding for at least 10 weeks duration. Using similar analytic techniques as described previously, we found that CHDPs did not have a significant effect on nonexclusive breastfeeding for at least 10 weeks (AOR = 0.85; 95% CI = 0.63, 1.14; Table 3). Factors associated with nonexclusive breastfeeding for at least 10 weeks duration included maternal age, race/ethnicity, education, and family income.
We found that almost two thirds of women who initiated breastfeeding in the hospital reported having been given commercial hospital discharge packs by hospital staff. Distribution of CHDPs gives new mothers a mixed message, because hospital staff may verbally discourage formula feeding, encourage initial attempts to breastfeed, and even instruct a woman on the proper technique of latching on. Reiff, for example, found that hospital “modeling” of the use of formula had greater influence on mothers than did verbal instruction that discouraged formula use.25
Since the early 1980s, there have been many studies, of widely varying quality and conclusions, of the impact of CHDPs on breastfeeding. The best of these studies compared receipt of discharge packs that contained formula with receipt of no discharge packs or of discharge packs without formula. Snell et al.,26 who studied 88 low-income Hispanic women in California, found that receipt of a gift pack that contained formula (compared with receipt of no gift pack) was associated with a statistically significant decrease in exclusive breastfeeding at 3 weeks. Frank et al.,27 who studied 343 low-income women in Boston, found that receipt of a gift pack that contained formula (compared with receipt of a gift pack without formula) was associated with a statistically significant decrease in exclusive breastfeeding at 4 months. By contrast, Evans et al.,28 Feinstein et al.,29 and Neifert et al.30 examined breastfeeding exclusivity and duration among women who were given a gift pack that contained formula compared with women who were given a gift pack that did not contain formula. In these 3 studies, no statistically significant differences were found with regard to breastfeeding duration among the study groups.
There have been several studies19,31,32 that compared receipt of hospital discharge packs that contained formula to receipt of packs that contained manual breast pumps, but these are difficult to interpret because breast pumps may be associated with increased breastfeeding.33 Another study34 compared receipt of a discharge pack that contained formula with receipt of no discharge pack and found a nonsignificant decrease in any breastfeeding, but they did not explore exclusive breastfeeding.
Our review of the literature led us to use exclusive breastfeeding as our primary outcome of interest. The PRAMS survey questions, however, allowed us to conduct statistical analyses of the responses on the effect of CHDPs on both exclusive and nonexclusive breastfeeding. We found that there was no association between nonexclusive breastfeeding for at least 10 weeks and the receipt of a CHDP. Other studies also have found no statistical association between nonexclusive breastfeeding and CHDPs.35 The reason for the lack of an association is not clear.
One limitation of our work is the self-report nature of our data. Recall bias is possible because women responded to our survey, on average, about 15 weeks after their infant’s birth. A recent review found that maternal recall of breastfeeding was valid and reliable in that time period,36 but there is no empirical evidence of whether women accurately remember having received a CHDP from the birthing hospital.
We may have underestimated the number of breastfeeding women who received formula promotional material. The PRAMS survey asked mothers whether they had received a “gift pack containing formula” from the birthing hospital. However, there were no questions about whether mothers had received formula manufacturers’ coupons or commercially produced literature on infant feeding in lieu of or in addition to the formula sample. This may have led us to underestimate the proportion of new mothers whose breastfeeding decisions were influenced by formula manufacturers’ inducements.
Our study is cross-sectional. Many previous studies of the association between discharge packs and breastfeeding were randomized controlled trials. Most were small studies done in urban academic medical centers; many included only low-income participants. The PRAMS survey methods accounted for the underrepresentation of certain sectors of the population (e.g., racial/ethnic minorities), and the data were weighted for nonresponse and noncoverage. However, we cannot say for certain that nonrespondents would have provided answers to PRAMS survey questions similar to those who responded to the survey. Causality cannot be established because the data reported are cross-sectional. Nevertheless, our study is population-based, drawn from a large, stratified, random sample of urban and rural Oregon women from several racial/ethnic populations who had a live birth.
The last limitation of our study is that it cannot be generalized beyond Oregon. According to the 2003 National Immunization Survey (the closest data available to our 2000–2001 birth cohort), more women were exclusively breastfeeding at 3 months in Oregon than in any other state.37
The production and sale of infant formula is big business. Although formula was originally produced for infants whose mothers could not nurse, formula is now marketed to almost all women. Formula manufacturers, endeavoring to increase sales, provide free formula to hospitals for in-hospital use in exchange for the opportunity to distribute formula samples to new mothers before they leave the hospital. Even women who have initiated breastfeeding in the hospital have become targets for formula manufacturers’ marketing and regularly receive CHDPs at the time of newborn hospital discharge. Some of these women may discontinue exclusive breastfeeding sooner than they would have without the marketing of formula. With rare exceptions, exclusive breastfeeding is the best form of infant feeding for the first 6 months of an infant’s life. This study indicates that provision of CHDPs to new mothers who have initiated breastfeeding may be associated with early discontinuation of exclusive breastfeeding. One way to increase exclusive breastfeeding may be to halt the provision of CHDPs at the time of newborn hospital discharge.
Notes. WD = weighted distribution; CHDP = commercial hospital discharge pack; WIC = Special Supplemental Nutrition Program for Women, Infants, and Children; BMI = body mass index. aMaternal prepregnancy body mass index (weight in kilograms divided by height in meters squared). Notes. OR = odds ratio; CI = confidence interval; CHDP = commercial hospital discharge pack; WIC = Special Supplemental Nutrition Program for Women, Infants, and Children; BMI = body mass index. aUnweighted number of survey respondents. bMaternal prepregnancy body mass index (weight in kilograms divided by height in meters squared). Notes. OR = odds ratio; CI = confidence interval; CHDP = commercial hospital discharge pack; WIC = Special Supplemental Nutrition Program for Women, Infants, and Children; BMI = body mass index. aUnweighted number of survey respondents. bMaternal prepregnancy body mass index (weight in kilograms divided by height in meters squared).
Exclusive Breastfeeding < 10 Weeks (n = 1598) Exclusive Breastfeeding ≥ 10 Weeks (n = 1086) Characteristic No. (%) WD % No. (%) WD % P Maternal age, y < 18 77 (4.8) 4.8 23 (2.1) 1.3 .007 18–34 1337 (83.7) 83.8 950 (87.5) 87.0 > 34 184 (11.5) 11.4 113 (10.4) 11.7 Race/ethnicity Hispanic 359 (22.5) 12.5 361 (33.2) 16.8 < .001 American Indian/Alaska Native 183 (11.5) 1.5 119 (11.0) 1.3 Asian/Pacific Islander 293 (18.3) 5.3 165 (15.2) 4.0 Non-Hispanic Black 168 (10.5) 1.9 77 (7.1) 1.1 Nn-Hispanic White 595 (37.2) 78.8 364 (33.5) 76.9 Education, y 0–8 114 (7.2) 5.6 129 (12.0) 6.2 < .001 9–11 238 (15.1) 14.2 162 (15.1) 10.6 12 523 (33.2) 36.5 257 (24.0) 25.3 > 12 700 (44.4) 43.7 525 (48.9) 58.0 Prepregnancy family income < $15 000 420 (28.6) 23.1 286 (28.3) 16.5 < .010 $15 000–$30 000 436 (29.6) 28.4 282 (27.9) 27.4 > $30 000 615 (41.8) 48.5 443 (43.8) 56.1 Marital status Not married 597 (37.4) 30.0 288 (26.5) 20.4 < .001 Married 1001 (62.6) 70.0 798 (73.5) 79.6 Smoking status Yes 283 (17.8) 20.2 71 (6.6) 6.9 < .001 No 1303 (82.2) 79.8 1004 (93.4) 93.1 Parity Primapara 763 (47.7) 47.1 464 (42.7) 38.8 .009 Multipara 835 (52.3) 52.9 622 (57.3) 61.2 CHDP Received 1177 (73.7) 69.2 700 (64.5) 61.6 .012 Did not receive 421 (26.3) 30.8 386 (35.5) 38.4 WIC client during pregnancy Yes 374 (42.8) 50.3 259 (42.5) 48.7 .091 No 500 (57.2) 49.7 351 (57.5) 51.3 BMIa < 25.0 889 (59.8) 60.4 607 (64.4) 65.9 .088 ≥ 25.0 597 (40.2) 39.6 336 (35.6) 34.1 Characteristic No.a Breastfed Child for < 10 Weeks, Weighted % Bivariate OR (95% CI) Multivariate OR (95% CI) Total 2684 57.1 Received CHDP Yes 1877 60.0 1.40 (1.08, 1.83) 1.39 (1.05, 1.84) No (Ref) 807 51.6 1.00 1.00 Age, y < 18 100 82.6 3.71 (1.64, 8.39) 2.89 (1.11, 7.52) 18–34 (Ref) 2287 56.2 1.00 1.00 > 34 297 56.6 1.02 (0.68, 1.51) 1.23 (0.80, 1.89) Race/ethnicity Hispanic 720 49.9 0.73 (0.59, 0.90) 0.53 (0.39, 0.72) American Indian/Alaska Native 302 61.3 1.16 (0.91, 1.48) 0.96 (0.73, 1.27) Asian/Pacific Islander 458 63.9 1.30 (1.02, 1.65) 1.47 (1.13, 1.90) Non-Hispanic Black 245 69.9 1.70 (1.28, 2.26) 1.57 (1.14, 2.17) Non-Hispanic White (Ref) 959 57.7 1.00 1.00 Education, y 0–8 243 54.3 1.19 (0.81, 1.75) 1.47 (0.89, 2.41) 9–11 400 64.1 1.79 (1.22, 2.63) 1.69 (1.03, 2.77) 12 780 65.6 1.91 (1.42, 2.58) 1.95 (1.39, 2.73) > 12 (Ref) 1225 49.9 1.00 1.00 Prepregnancy family income < $15 000 706 64.3 1.62 (1.19, 2.22) 1.49 (1.01, 2.21) $15 000–$30 000 718 57.1 1.20 (0.88, 1.62) 1.11 (0.79, 1.57) > $30 000 (Ref) 1058 52.6 1.00 1.00 Marital status Not married 885 66.3 1.68 (1.26, 2.23) Married (Ref) 1799 53.9 1.00 Smoking status Yes 354 79.5 3.41 (2.17, 5.36) No (Ref) 2307 53.3 1.00 Parity Primapara 1227 61.8 1.40 (1.09, 1.80) Multipara (Ref) 1457 53.5 1.00 WIC client Yes (Ref) 1230 60.6 1.00 No 1247 59.1 0.80 (0.62, 1.04) BMIb < 25.0 (Ref) 1496 55.6 1.00 ≥ 25.0 933 61.3 1.27 (0.97, 1.67) Characteristic No.a Breastfed Child for < 10 Weeks, Weighted % Bivariate OR (95% CI) Multivariate OR (95% CI) Total 3280 25.5 Received CHDP Yes 2320 24.8 0.89 (0.67, 1.17) 0.85 (0.63, 1.14) No (Ref) 960 27.1 1.00 1.00 Age, y < 18 years 136 48.9 2.89 (1.60, 5.21) 1.81 (0.86, 3.81) 18–34 years (Ref) 2812 24.9 1.00 1.00 > 34 years 332 22.1 0.86 (0.55, 1.34) 1.27 (0.78, 2.09) Race/ethnicity Hispanic 958 23.4 0.88 (0.71, 1.10) 0.55 (0.39, 0.78) American Indian/Alaska Native 370 31.4 1.32 (1.03, 1.68) 1.04 (0.78, 1.39) Asian/Pacific Islander 538 22.7 0.85 (0.66, 1.09) 0.94 (0.71, 1.26) Non-Hispanic Black 342 37.2 1.71 (1.32, 2.21) 1.39 (1.03, 1.88) Non-Hispanic White (Ref) 1072 25.8 1.00 1.00 Education, y 0–8 324 27.8 1.77 (1.12, 2.79) 1.83 (1.01, 3.31) 9–11 532 34.0 2.37 (1.61, 3.48) 1.53 (0.92, 2.56) 12 976 31.9 2.16 (1.58, 2.95) 1.90 (1.35, 2.67) > 12 (Ref) 1396 17.8 1.00 1.00 Prepregnancy family income < $15 000 911 34.9 2.39 (1.71, 3.33) 2.31 (1.55, 3.46) $15 000–$30 000 905 27.1 1.66 (1.19, 2.31) 1.60 (1.10, 2.33) > $30 000 (Ref) 1175 18.3 1.00 1.00 Marital status Not married 1160 38.5 2.42 (1.85, 3.17) Married (Ref) 2120 20.5 1.00 Smoking status Yes 453 49.5 3.62 (2.54, 5.15) No (Ref) 2799 21.3 1.00 Parity Primapara 1452 27.1 1.16 (0.89, 1.50) Multipara (Ref) 1828 24.3 1.00 WIC client Yes (Ref) 1612 31.0 1.00 No 1443 22.1 0.63 (0.48, 0.82) BMIb < 25.0 (Ref) 1812 24.1 1.00 ≥ 25.0 1132 29.0 1.29 (0.97, 1.70)
The authors thank the Maternal and Child Health Bureau of the US Department of Health and Human Services and the Centers for Disease Control and Prevention for their financial and technical support of the Oregon Pregnancy Risk Assessment Monitoring System.
We also thank Tina Kent for her work with the Oregon Pregnancy Risk Assessment Monitoring System and Michelle Adler, Thomas Brundage, Andrea C. Kovach, Lawrence Grummer-Strawm, Charlotte McKay, Anne Merewood, Andy Osborn, and Marcia Walker for their contributions.
Human Participant Protection The Oregon Pregnancy Risk Assessment Monitoring System study protocols and informed consent procedures were approved by the Oregon State Public Health/Multnomah County Public Health institutional review board.
