Effect of single-dose anthelmintic treatment during pregnancy on an infant’s response to immunisation and on susceptibility to infectious diseases in infancy: A randomised, double-blind, placebo-controlled trial

Helminth infections affect the human immune response. We investigated whether prenatal exposure to and treatment of maternal helminth infections affects development of an infant’s immune response to immunisations and unrelated infections. In this randomised, double-blind, placebo-controlled trial, we enrolled 2507 women in the second or third trimester of pregnancy who were planning to deliver in Entebbe General Hospital, Entebbe, Uganda. With a computer-generated random number sequence in blocks of 100, we assigned patients to 440 mg albendazole and 40 mg/kg praziquantel (n=628), 440 mg albendazole and a praziquantel-matching placebo (n=625), 40 mg/kg praziquantel and an albendazole-matching placebo (n=626), or an albendazole-matching placebo and praziquantel-matching placebo (n=628). All participants and hospital staff were masked to allocation. Primary outcomes were immune response at age 1 year to BCG, tetanus, and measles immunisation; incidence of infectious diseases during infancy; and vertical HIV transmission. Analysis was by intention-to-treat. This trial is registered, number ISRCTN32849447. Data were available at delivery for 2356 women, with 2345 livebirths; 2115 (90) of liveborn infants remained in follow-up at 1 year of age. Neither albendazole nor praziquantel treatments affected infant response to BCG, tetanus, or measles immunisation. However, in infants of mothers with hookworm infection, albendazole treatment reduced interleukin-5 (geometric mean ratio 0·50, 95 CI 0·30-0·81, interaction p=0·02) and interleukin-13 (0·52, 0·34- 0·82, 0·0005) response to tetanus toxoid. The rate per 100 person-years of malaria was 40·9 (95 CI 38·3-43·7), of diarrhoea was 134·1 (129·2-139·2), and of pneumonia was 22·3 (20·4-24·4). We noted no effect on infectious disease incidence for albendazole treatment (malaria [hazard ratio 0·95, 95 CI 0·79-1.14], diarrhoea [1·06, 0·96-1·16], pneumonia [1·11, 0·90-1·38]) or praziquantel treatment (malaria [1·00, 0·84-1·20], diarrhoea [1·07, 0·98-1·18], pneumonia [1·00, 0·80-1·24]). In HIV-exposed infants, 39 (18) were infected at 6 weeks; vertical transmission was not associated with albendazole (odds ratio 0·70, 95 CI 0·35-1·42) or praziquantel (0·60, 0·29-1·23) treatment. These results do not accord with the recently advocated policy of routine antenatal anthelmintic treatment, and the value of such a policy may need to be reviewed. Wellcome Trust. © 2011 Elsevier Ltd. The study is described in detail elsewhere.18 Briefly, the study area was Entebbe Municipality and Katabi subcounty, beside Lake Victoria, Uganda. The area is occupied by urban, rural, and fishing communities. Helminth infection is highly prevalent in the area,19 and malaria, diarrhoea, and pneumonia are common in young children.20 The study population consisted of pregnant women who presented at the government-funded antenatal clinic at Entebbe General Hospital between April 9, 2003, and Nov 24, 2005, where roughly 70% of pregnant women from the study area received antenatal care.21 Women were included if resident in the study area, planning to deliver in the hospital, willing to know their HIV status, and in the second or third trimester of pregnancy. They were excluded if they had possible helminth-induced pathological changes (haemoglobin <80 g/L, clinically apparent severe liver disease, or diarrhoea with blood in stool), a history of an adverse reaction to anthelmintics, already been enrolled in the trial during an earlier pregnancy, or if the pregnancy was deemed abnormal by a midwife. All participants gave written, informed consent. Ethics approval was given by the Uganda Virus Research Institute, Uganda National Council for Science and Technology, and London School of Hygiene and Tropical Medicine. We used a two-by-two factorial design to randomly assign patients in a 1:1:1:1 ratio to receive simultaneously either single-dose albendazole (440 mg) and single-dose praziquantel (40 mg/kg), albendazole and a praziquantel-matching placebo, an albendazole-matching placebo and praziquantel, or an albendazole-matching placebo and a praziquantel-matching placebo (albendazole and matching placebo, Glaxosmithkline, Brentford, UK; praziquantel and matching placebo, Medochemie Ltd, Limassal, Cyprus). The randomisation code was generated by the trial statistician with a computer-generated random number sequence, with block size 100. Treatments were packed in sealed envelopes and labelled with an allocation number by colleagues at the Medical Research Council Unit in Entebbe who did not otherwise contribute to the trial. Treatments were allocated in numerical order by trained interviewer-counsellors who observed the patients taking the treatment correctly on enrolment to the study. Treatment allocation was masked from all participants and staff during the study. Demographic and clinical details, and blood samples were obtained at screening; stool samples were obtained before enrolment. After enrolment, women continued to receive standard antenatal care, including haematinics, tetanus immunisation, and intermittent presumptive treatment for malaria twice after their first trimester of pregnancy; women with HIV were offered intrapartum and neonatal single-dose nevirapine for prevention of mother-to-child (vertical) HIV transmission.22 Stool samples were obtained after delivery to assess effectiveness of anthelmintic treatment; thereafter, all mothers received praziquantel and albendazole. Infants received BCG and polio immunisation at birth; diphtheria, pertussis, tetanus, Haemophilus influenzae, hepatitis B, and polio immunisation at 6, 10, and 14 weeks; and measles immunisation at 9 months. Children also attended the study clinic when unwell; doctors diagnosed, treated, and recorded their illnesses. Community fieldworkers visited each participant's home twice a month, measured the child's temperature, and recorded symptoms reported by the child's carer. At age 12 months, blood and stool samples were obtained from the children and growth outcomes were measured. Children who were unwell at their 12-month visit were given appropriate treatment and asked to return to complete the visit procedures when well. The primary outcomes were immune response at age 1 year to BCG, tetanus, and measles immunisation; incidence of malaria, diarrhoea, pneumonia, measles, and tuberculosis during infancy as diagnosed by doctors at the study clinic; and vertical HIV transmission.18 Planned secondary outcomes were growth and anaemia at age 1 year. Community-reported data for illness events were included as a secondary outcome for comparison with doctor-diagnosed illness events from clinic visits. We also considered two additional unplanned secondary outcomes: infant mortality and asymptomatic malaria (presence of malaria parasitaemia) at 1 year of age. Cytokine responses at 1 year of age to crude culture filtrate proteins of Mycobacterium tuberculosis (cCFP) were measured as an indicator of response to BCG immunisation. Cytokine responses at age 1 year to tetanus toxoid were measured as an indicator of response to tetanus immunisation. We examined stimulated interferon-γ (type 1), interleukin-5 (type 2), interleukin-13 (type 2), and interleukin-10 (regulatory) responses in a whole-blood assay, as previously described.23 Total serum IgG, IgG4, and IgE responses to tetanus toxoid were measured by ELISA (webappendix p 1). Total serum measles-specific IgG was measured by ELISA (Dade Behring/Siemens, Eschborn, Germany) according to the manufacturer's protocol. Immunological assays were done after all samples had been obtained, in a randomised sequence (by use of a computer-generated random number sequence), to avoid confounding of secular trends with variations in assay performance. For the primary outcome, doctor-diagnosed illness events, clinical malaria was fever (temperature ≥37·5°C) with parasitaemia; diarrhoea was an infant's carer's definition, with stool frequency recorded;24 pneumonia was cough with difficulty in breathing, and age-specific fast breathing;25 measles was defined by standard clinical criteria and confirmed by measurement of specific antibody;26 and children with suspected tuberculosis were investigated as clinically indicated.27 For the secondary outcome, community-reported illnesses, febrile illness was defined as measured by fieldworkers (temperature ≥37·5°C) or as reported by the child's carer; diarrhoea as reported by the carer, with stool frequency recorded; presumptive pneumonia was cough with difficulty in breathing, or age-specific fast breathing as measured by fieldworkers. Stool samples were examined for helminth ova with the Kato-Katz method28 and by charcoal culture for Strongyloides sterocoralis infection;29 two Kato-Katz slides were prepared from each sample and examined for hookworm ova within 30 mins of preparation, or examined the next day for other species. Hookworm and Schistosoma mansoni infections were classified into low, medium, and high intensities according to WHO guidelines.30 Blood samples were examined by a modified Knott's method for Mansonella perstans31 and by thick film for malaria parasites. Haemoglobin was estimated by Coulter analyser (Beckman Coulter, Nyon, Switzerland). Quality control for Kato-Katz analyses was provided by the Vector Control Programme of the Ministry of Health, Uganda, and for haematology and malaria parasitology through the UK National External Quality Assessment Schemes. Mothers' HIV serology was done by rapid test algorithm.30 Blood was obtained from cord and at 6 weeks of age from infants of mothers with HIV for assessment of vertical HIV transmission. Plasma and whole blood cell pellet were separated by centrifugation and stored at −80°C until assays were done. For detection of HIV-1 proviral DNA in infants at 6 weeks, DNA was extracted from stored whole blood cell pellets and amplified by nested PCR of three conserved viral regions, tat, gp41, and nef (webappendix p 1). For both cord and 6-week samples, plasma HIV load was measured with Bayer Versant branched DNA assay version 3.0 (Bayer, Leverkusen, Germany) or Roche Amplicor HIV-1 RNA Monitor test version 1.5 (Roche, NJ, USA). Infants were regarded as being HIV positive if the 6-week sample had a positive DNA PCR for any of the viral regions and a viral load of 1000 copies per mL or more; for four infants, only viral load data were available, so they were used to establish HIV status. Viral load and DNA-PCR results were concordant apart from one infant (viral load 6699 copies per mL, PCR negative) who was seronegative by rapid test algorithm at age 18 months and was classified as HIV negative. In infants with HIV infection, transmission was regarded as likely to have been intrauterine if the viral load in cord blood was 1000 copies per mL or more. Analysis was done after all children were older than 15 months. Data for samples and measurements obtained at routine, 1-year visits were included if the child attended within 2 months after their first birthday. Data for illness events and mortality were censored strictly at 1 year. Results for younger twins were excluded from all analyses. On the basis of our preliminary study,32 the planned cohort size of 2500 was expected to accrue 1860 person-years of follow-up in infancy and 1594 infants were expected to be seen at age 1 year. For either maternal treatment, assuming no interaction between treatments, this number would give 80% power to detect rate ratios of 0·82 for malaria, 0·91 for diarrhoea, and 0·76 for pneumonia, with p values of less than 0·05, assuming frequency of disease in the placebo groups to be 50 per 100 person-years for malaria,33 190 per 100 person-years for diarrhoea,34 and 25 per 100 person-years for pneumonia.35 Incidence of both tuberculosis and measles was expected to be low,18 therefore only very large differences in incidence would be detected. Samples from 1594 infants assessed at 1 year would detect differences in infant cytokine responses of 0·11 log10 between intervention groups.18,32 The patients that were included in analysis for each of the primary outcomes differed because data for each outcome was obtained at different times. For immune response at age 1 year, analysis included all children who provided a blood sample at 1 year and who had received full BCG (for cCFP analysis) or tetanus (for tetanus toxoid analysis) immunisation at Entebbe Hospital; second-born twins were excluded. For incidence of infectious diseases during infancy, analysis included all liveborn children, excluding second-born twins. For vertical HIV transmission, analysis included all children whose mothers were not receiving highly-active antiretroviral therapy, and from whom blood samples at age 6 weeks were available; second-born twins were excluded. Cytokine and antibody responses showed skewed distributions, with disproportionate numbers of zero values. Results were transformed to log10(concentration+1) and analysed by linear regression with bootstrapping to estimate bias-corrected accelerated confidence intervals.36 Regression coefficients were back-transformed to give geometric mean ratios. Interactions were examined with Wald tests. For doctor-diagnosed disease incidence, time at risk began at birth and was censored at loss to follow-up, death, or age 1 year. All children with known date of birth were included in the analysis until censoring, irrespective of whether they had made a clinic visit for illness. For each disease, we calculated incidence rates for all events. Disease episodes within 14 days of an initial presentation with the same disease were regarded as part of the same episode and excluded from the analysis; time at risk was adjusted accordingly, excluding these 14-day periods from the total person-time denominator. Hazard ratios (HRs) for effects of treatment were calculated with Cox regression, with robust SEs to allow for within-child clustering. For community-reported illness data, a generalised-estimating-equation approach with exchangeable correlation structure was used to model effects of treatment on repeated binary outcomes. Odds ratios (ORs) for the effects of treatment on vertical HIV transmission were calculated with logistic regression. The prevalence of asymptomatic malaria at 1 year was compared between treatment groups with logistic regression. Infant mortality per 1000 livebirths was estimated from Kaplan-Meier survival probabilities to age 1 year, and effects of maternal anthelmintic treatment were assessed by Cox regression. Weight-for-age, height-for-age, and weight-for-height Z scores at 1 year were derived from WHO growth standard reference scales, with igrowup macros. We examined effects of maternal treatment on Z scores and on haemoglobin at 1 year by linear regression. We did two prespecified subgroup analyses, examining effects of albendazole treatment in children of mothers with a hookworm infection, and effects of praziquantel treatment in children of mothers with schistosomiasis. Differences between subgroups were examined by fitting interaction terms in regression models. All p values were two-sided with no adjustment made for multiple comparisons. Analyses were done with Stata 10.1. The sponsor of the study had no role in study design, data collection, data analysis, data interpretation, or writing of the report. AME had full access to all the data in the study and had final responsibility for the decision to submit for publication.