Uk Deborah Manchester

Baker

Dr Adam Garrow
email: a.garrow@salford.ac.uk

Christopher Shiels

INEQUALITIES IN IMMUNISATION AND BREASTFEEDING IN AN ETHNICALLY DIVERSE URBAN AREA: CROSS SECTIONAL STUDY IN

Objectives. To examine inequalities in immunisation and breastfeeding by ethnic group and their relation to relative deprivation. Design. Cross -sectional study. Setting. Manchester, UK Participants. 20,203 children born in Manchester (2002-7), who had been coded as of White, Mixed, Indian, Pakistani, Bangladeshi and Black or Black British ethnicity in the Child Health System database. Main Outcome Measures. Breastfeeding at 2 weeks postpartum; uptake of Triple Vaccine (Diptheria, Pertussis and Tetanus) at 16 weeks postpartum; uptake of the Measles, Mumps and Rubella Vaccine (MMR) by the age of 2.

Results. Black or Black British infants had the highest rates of breastfeeding at 2 weeks postpartum (89%), and South Asian infants had highest Triple and MMR vaccination rates (Indian, 95%, 96%; Pakistani 95%, 95%; Bangladeshi 96.%, 95%) after controlling for area level of deprivation, parity, parenthood status and age. White infants were least likely to be breastfed at 2 weeks postpartum (36%), and to be vaccinated with Triple (92%) and MMR vaccines (88%). Within the White ethnic group lower percentages of immunisation and breastfeeding were significantly associated with living in a deprived area and with increasing parity. This was not found within Black or Black British and Pakistani ethnic groups.

Discussion Practices that are protective of child health were consistently less likely to be adopted by White mothers living in deprived areas. Methods of health education and service delivery that are designed for the general population are unlikely to be successful in this context and evidence of effective interventions needs to be established.

INTRODUCTION

The uptake and adoption of preventative interventions by mothers of young children is a key component of public health policy to reduce child health inequalities in the UK. [START_REF]Every Child Matters: Change for Children in Health Services[END_REF]2] Immunisation and breastfeeding provide children with a healthy start in life, not least by protecting against infectious diseases; [3, 4] increasing rates of breastfeeding in disadvantaged groups is identified as an important step in actions to reduce inequalities in infant mortality. [START_REF]Review of the Infant Mortality Health Inequalities PSA Target[END_REF] Key social and demographic indicators of disadvantage that are associated with lower uptake of immunisation and/or breastfeeding include lower socioeconomic status, lower level of education, teenage motherhood, single parenthood and multi parity. [6][START_REF] Pearce | Factors associated with uptake of measles, mumps, and rubella vaccine (MMR) and use of single antigen vaccines in a contemporary UK cohort: prospective cohort study[END_REF][START_REF] Griffiths | The contribution of parental and community ethnicity to breastfeeding practices: evidence from the Millennium Cohort Study[END_REF] Delivering interventions that have an impact on inequalities in practices that are protective of child health is a particular challenge in inner city areas of the UK with ethnically diverse populations, since Black and Ethnic Minority (BME) groups tend to be clustered in the most deprived neighbourhoods. [START_REF]Census: Key Statistics for Local Authorities[END_REF] The extent to which ethnic group determines inequalities over and above those associated with poverty is thus fundamental to the equitable provision of preventative services in infancy and early childhood.

Influential evidence on the determinants of inequalities in health by ethnic group in adulthood has, in general, shown that the relationship between poorer health and membership of a BME group is explained by relative deprivation, given the concentration of BME groups in deprived areas and thus the effects of poverty and social disadvantage on their health. [START_REF] Nazroo | Genetic, cultural or socio-economic vulnerability? Explaining ethnic inequalities in health[END_REF]11] Infant mortality rates (IMR) are currently used as the headline indicator to measure child health inequality in the UK and these suggest some degree of health disadvantage in infancy associated with BME status. [START_REF] Miles | Large differences in infant mortality by ethnic group[END_REF] In 2005 IMRs in both Pakistani and Caribbean ethnic groups were twice that of the White British group. However the extent to which IMR constitutes a reliable marker of child health inequality in this context is questionable for two reasons. Firstly evidence suggests that high rates of infant mortality for women of Pakistani origin are associated with congenital anomalies and those for Caribbean women with very preterm births. These conditions are unlikely to be related to deprivation. [START_REF] Bakeo | Investigating variations in infant mortality in England and Wales by mother's country of birth, 1983-2001[END_REF] Secondly there are a relatively small number of infant deaths in this country. The rate of infant deaths was 5.00 per 1000 population in England in 2002-4 [START_REF]Review of the Infant Mortality Health Inequalities PSA Target[END_REF] ranging from 1.6 per 1000 live births to 11.9 per 1000 live births in the 354 local authority areas.

Measurement problems with using small numbers are compounded when breakdown of the figures by ethnicity are considered, since BME groups constitute a relatively small proportion of the UK population (8% according to the 2001 Census).

In this paper we examine the extent to which deprivation is the common factor that explains inequalities both between and within ethnic groups across three contrasting preventative practices in childhood, uptake of Measles, Mumps and Rubella (MMR) vaccine, uptake of the Triple vaccine (Diptheria, Pertussis and Tetanus) and breast feeding. In this endeavour we are focusing not directly on health itself, but on maternal behaviours that are health promoting. The relation between deprivation and adoption of such behaviours is likely to be mediated by factors as various as access to health care, beliefs about the efficacy of an intervention, cultural /social group norms and support networks. Interestingly, recent studies that focus on only one of these three preventative practices, demonstrate higher rates of immunisation and breastfeeding for South Asian when compared with White infants, even when the confounding effect of relative deprivation has been taken into account, although there is less consistency for Black or Black British infants; some studies have found lower rates of immunisation for this group, compared with South Asian and White infants, although these have not consistently controlled for relative deprivation. [START_REF] Pearce | Factors associated with uptake of measles, mumps, and rubella vaccine (MMR) and use of single antigen vaccines in a contemporary UK cohort: prospective cohort study[END_REF][START_REF] Griffiths | The contribution of parental and community ethnicity to breastfeeding practices: evidence from the Millennium Cohort Study[END_REF][14][START_REF] Hawker | Widening inequalities in MMR vaccine uptake rates among ethnic groups in an urban area of the UK during a period of vaccine controversy (1994-2000)[END_REF][START_REF] Mixer | Ethnicity as a correlate of the uptake of the first dose of mumps, measles and rubella vaccine[END_REF][START_REF] Samad | Differences in risk factors for partial and no immunisation in the first year of life: prospective cohort study[END_REF][START_REF] Baker | Differences in rate of uptake of immunisation among ethnic groups[END_REF] The study presented here builds on this evidence by comparing immunisation and breast feeding practice between White, Mixed, South Asian and Black or Black British ethnic groups for infants born in the city of Manchester, the 4 th most deprived local authority in the UK, where in some areas BME groups constitute over one third of the population. In addition we examine the socioeconomic and demographic factors that are associated with variations in these practices within the three largest ethnic groups in this area -Black or Black British, Pakistani and White. No previous research has undertaken such a comparative analysis for a community based population.

METHODOLOGY

Data were drawn from the Child Health System database for the city of Manchester UK. These data were collected from two sources: 1. The formal notification of birth which contains mandatory information collected by the midwife immediately after the birth of all babies and includes basic demographic details of mother and child, previous obstetric history and birth details. 2. A Personal Child Health Record that is issued on the birth of every child by all the Manchester Primary Care Trusts. This document is the main record of the child's health, growth and development and is used by everyone involved in the care of the child from birth to school entry and contains information about feeding practice and child immunisations from 8 weeks to 3-5 years. Each section of the Child Health Record is produced in triplicate with the top copy remaining in the Child Health Record and on completion, one copy is retained by the Health Visitor and the third copy returned to The Child Health Department for entry onto the database.

This study covers the period from 1/4/2002 and 31/3/2007 during which a total of 31521 births were recorded. However, this analysis is limited to 20,203 children born during the study period coded as being of White, Mixed, Indian, Pakistani, Bangladeshi and Black or Black British ethnicity in the Child Health System database. Together, these ethnic groupings make up approximately 97% of the city of Manchester population. The following variables were extracted from this database:-

Maternal Characteristics

Self report of ethnicity has been collected on this database since 2002, using a categorisation based on the 2001 Census. The following are the ethnic group categories that were used as the basis for analysis by ethnic group in this study:

• White: this category includes White British, White Irish and any other White background.

• Mixed: this category includes White and Black Caribbean, White and Black African, White and Asian and any other Mixed background (e.g. Black and Asian, Black and Chinese, Black and White, Chinese and White, Asian and Chinese).

• Asian and Asian British. Data for Indian, Pakistani and Bangladeshi (South Asian ethnic groups) have been included from this category.

• Black or Black British. This category includes Black Caribbean, Black African, and any other Black background.

Other maternal characteristics that have been used in this analysis are age (<17, 17-18, 19-25, 26-29 30-34, 35+) parenthood status (one parent family or not) and parity (no previous births, 1 previous birth, 2 previous births, 3 or more previous births).

The measurement of breastfeeding and immunisation

Breastfeeding at two weeks postpartum is recorded as 'fully breastfed', 'partially breastfed' or 'not breastfed'. For this analysis the first two categories have been combined to produce the dichotomous variable 'breastfed : yes/no', since the intention of the analysis was to establish the uptake of breastfeeding as a health promoting behaviour and its variation by ethnic group. The Triple Vaccine immunisation against Diptheria, Pertussis and Tetanus is offered to children 8, 12 and 16 weeks after birth.

The Child Health System database records information on the children who have received all three inoculations as a single variable and this has been used to measure uptake ('yes/no'). The Measles, Mumps and Rubella vaccine (MMR) is offered to children at 13 months, but can be administered up to the age of 2; this was the cut-off point for measurement of uptake.

The measurement of deprivation

Area level of deprivation was measured using the Income Deprivation Affecting Children (IDAC) Index that is a sub-set of the Income Deprivation domain of the English Index of Multiple Deprivation 2004. [START_REF]The English Indices of Deprivation[END_REF] The calculation of the IDAC index is based on the percentage of children under 16 who are living in families in receipt of Income Support and Income-Based Job Seekers Allowance or in families in receipt of Workers Families Tax Credit or Disabled Persons Tax Credit whose equivalised income is below 60% of the median, before housing costs.

Using IDAC scores for the constituent Lower Super Output Areas, the mean level of area deprivation for each of the 32 wards in Manchester was calculated and divided into quartiles. The most affluent wards in Manchester had a range of IDAC scores between 0.12 and <0.37; the 2nd quartile between >0.37 and <0.47 the 3rd quartile between >0.47 and < 0.54 and the least affluent quartile between >0.54 and 0.73. In the statistical analysis the 2 nd and 3 rd quartiles have been merged.

Data were anonymised at source. The Local Ethnics Committee agreed that in this circumstance ethics approval was not required for the study.

ANALYSIS

We first calculated the distribution of our sample by maternal age, parity, parenthood status, ethnic group and area level of deprivation using descriptive statistics. Logistic regression models were then used to examine the independent association of ethnic group with MMR uptake by 2yrs, Triple vaccine uptake by 16 weeks postpartum and breastfeeding at 2 weeks postpartum. The final analysis examined predictors of immunisation and breastfeeding within the main ethnic groups in Manchester -White, Pakistani and Black or Black British ethnic groups. The independent variables entered into logistic regression models were maternal age, parenthood status, parity and area deprivation level. Each of these variables was controlled for the others in the logistic regression models.

RESULTS

Table 1 shows that just under half of the sample lived in the most deprived areas of Manchester (43%). In addition 13% of the population were single parents, 1% were aged under 17 and 5% were aged between 17 and 18 years at the time of birth. The largest ethnic minority groups were Pakistani (15%) and Black or Black British (15%), with Whites constituting 51% of the sample.

Socioeconomic and demographic characteristics were differentially distributed across ethnic groups. 43% of White mothers and 51% of mothers with Mixed ethnicity lived in areas of high deprivation. But those most likely to live in these areas were of Black or Black British ethnicity (68%). In contrast, members of South Asian ethnic groups were more likely to live in more affluent areas. For example only 19% of Pakistanis lived in deprived areas. Single parenthood and teenage motherhood were most commonly characteristic of mothers of White (17%, 9% respectively), Mixed (18%,9%) and Black or Black British ethnicity (14%,5%), but were uncommon in South Asian ethnic minority groups. In terms of parity, members of Black or Black British, Pakistani and Bangladeshi ethnic groups were more likely to have larger families with two or more previous births (38%, 35%, 41% respectively) compared to White mothers (25%).

11

For health promoting behaviours, in the sample as a whole 90% of children had been MMR vaccinated, 93% had received all three courses of the Triple vaccine and 58%

were breastfed at 2 weeks postpartum. Table 2 shows that there were distinct patterns of MMR/Triple uptake and breastfeeding for the largest ethnic groups. White babies were least likely to have had the MMR vaccine (88%) and infants from Indian, Pakistani and Bangladeshi ethnic groups were most likely to have been vaccinated (96%, 95%, 95%).

The lowest uptake rate for the Triple vaccine was observed for White (92%) and Mixed ethnicity (91%) and highest for Indian (94%), Pakistani (95%) and Bangladeshi (96%) ethnic groups.

Breastfeeding was most common amongst Black or Black British mothers (89%).

Over two thirds of women in South Asian ethnic groups were breastfeeding at two weeks postpartum; the highest percentage was for Indians (85%). In contrast only 36% of White mothers were breastfeeding at this time, although this figure was higher for women of Mixed ethnicity (58%). Table 3 shows that for White mothers, living in a deprived area was significantly associated with lower uptake of the MMR and Triple vaccines and breastfeeding at two weeks postpartum. There were no significant relationships between these outcome measures and area level of deprivation for Pakistani mothers. Black or Black British mothers were significantly more likely to be breastfeeding at two weeks postpartum, if they lived in a deprived area (OR=2.41, 95% CIs= 1.51-3.86, p<0.001).

Multiparity was also a strong independent predictor of lower uptake across all three measures for White women only, with the likelihood of MMR and Triple vaccination and breastfeeding decreasing with increasing parity.

White women who were single parents were significantly less likely to breast feed than those with partners (OR=0.58, 95% CIs=0.46-0.74, p<0.001). The latter was also the case for Black or Black British mothers (OR=0.58, 95% CIs=0.35-0.89, p<0.02).

Younger White mothers, particularly those aged 25 or under, were significantly less likely to take up the Triple vaccine and to breastfeed at two weeks post partum when compared with older mothers (>35yrs). There was no similarly consistent relationship between age and uptake of vaccination for Pakistani and Black or Black British mothers.

DISCUSSION

Principal Findings

The health disadvantage of White infants living in deprived areas This research shows that ethnicity is associated with important variations in the practice of health promoting behaviours in infancy and that differences between ethnic groups remain after area level of deprivation has been taken into account. In this study Black or Black British infants were most likely to be breastfed and Pakistani and Bangladeshi infants most likely to be vaccinated with both MMR and Triple vaccines. Immunisation and breastfeeding were lowest for White infants.

This study exposes the unique disadvantage of children born to White mothers living in deprived areas. These mothers were more likely to be single parents or teenage mothers -both of which are indicators of social exclusion -but infants were also less likely to be breastfed and to be vaccinated. Typically uptake of immunisation and breastfeeding initiation became less likely with each new child that was born into a family. The health impact of socioeconomic disadvantage is thus compounded for White infants by lower adherence to practices conferring a health advantage. Most notably, this is not the case for Black or Black British infants whose chance of being brought up in a deprived area was even greater than that of the White infants in this sample. Within this ethnic group there were no significant associations between lower rates of breastfeeding and immunisation and area level of deprivation; this was also the case for Pakistanis, although these mothers were less likely to live in deprived areas.

Strengths and Weaknesses of Study Design

One of the main strengths of the study is its location in an inner city area with an ethnically diverse population. This enabled a breakdown of the BME sample into subgroups with substantial numbers that allowed for the analysis of inequalities within -as well as between ethnic groups. In order to maximise numbers for within group analyses, Black or Black British ethnicity was not disaggregated into its constituent ethnic groups. In this sample the majority of people in this category were Black African (64%), 19% were Black Caribbean and the remaining 17% were from 'any other Black background'. There were no significant differences in immunisation uptake between these groups although Black African women were significantly more likely to breastfeed (92% breastfeeding at two weeks postpartum) compared with Black Caribbean women (79% breastfeeding at two weeks postpartum).

The location also enabled a test for the confounding effect of deprivation in an area which is the 4 th most deprived local authority in England.

Access to routinely collected statistics on immunisation and breastfeeding enabled a comparison of patterns of uptake by ethnic group across three outcome measures.

Previous studies have focused on single outcomes only.

The poor coding of ethnic group in routinely collected data sources is a common problem for research in the UK. [START_REF] Moser | Assessing quality of NHS Numbers for Babies data and providing gestational age statistics[END_REF][START_REF] Mixer | Ethnicity as a correlate of the uptake of the first dose of mumps, measles and rubella vaccine[END_REF][START_REF] Hawker | Widening inequalities in MMR vaccine uptake rates among ethnic groups in an urban area of the UK during a period of vaccine controversy (1994-2000)[END_REF] In the time period covered by this study (2002 -2007) almost a third of entries on the Child Health System database were incomplete for ethnicity (9846/31521). Analyses of possible bias indicated that a large proportion of the missing values emanated from an area of Manchester with a predominantly White population and with no significant movement of BME groups into the area between the 1991 and 2001 Census. When the investigators compared the distribution of the sample by ethnicity with 2001 Census data for Manchester for children aged 0-4 it was clear that the missing values led to an under estimation of White ethnicity as a percentage of the total population and an overestimation of BME groups, particularly Black or Black British infants (Supplementary Table 1), although it is also a possibility that the sample in the Child Health System database is indicative of a trend towards the decline of the White population in ethnically diverse inner city areas and the increase of ethnic minority populations.[21] Following the assumption that the missing values were predominantly of White ethnicity did not significantly alter the demographic profile of the sample White population in terms of gender, age or parity (Supplementary Table 2). However it did indicate that those with missing ethnicity codes were less likely to be deprived and more likely to breastfeed than the White study sample. Including these values as 'White' would decrease the % living in deprived areas from 43% to 39% and increase those breastfeeding by 5%.

Entry of breastfeeding data onto the Child Health System database was also poor, with only 8946/20203 records for breastfeeding at 2 weeks postpartum. However the reliability of the findings are enhanced by replicating the results at another time point, for breastfeeding at initiation ( Supplementary Table 3). Tests for representativeness revealed that there were no significant differences by ethnic group or by area of deprivation between the characteristics of the whole sample and the missing values sample for breastfeeding at two weeks postpartum. It is therefore unlikely that missing values were a source of bias in the study. Moreover the numbers available for analyses of breastfeeding by ethnic group are more substantial than those of comparable studies based on the UK populations, in which breastfeeding data are based on retrospective self report. [START_REF] Griffiths | The contribution of parental and community ethnicity to breastfeeding practices: evidence from the Millennium Cohort Study[END_REF][START_REF] Kelly | Racial/ethnic differences in breastfeeding initiation and continuation in the United Kingdom and comparison with findings in the United States[END_REF] A further limitation of this study is that deprivation was measured at the area level rather than using individual level measures of socio economic or educational statuses, which were not available on the Child Health System database. Some proxies of social disadvantage at the individual level such as teenage motherhood and single parenthood were however available and these display similar trends to findings based on area level of deprivation.

Interpretation of results

Socioeconomic and cultural context of health promoting behaviours in infancy

The fact that patterns of association between immunisation and breastfeeding and relative deprivation within the White ethnic group do not generalise to Black or Black British and Pakistani ethnic groups suggests that the role of relative deprivation as a determinant of such behaviours is dependent on the extent to which other possible explanatory factors such as access to health care, beliefs about the efficacy of immunisation and breastfeeding, cultural/social norms and support networks vary according to socioeconomic status. For example a qualitative study that explored decisions about infant feeding method for mothers living in deprived areas in the UK found a characteristic set of norms and values associated with breastfeeding that were embedded in support networks. [START_REF] Bailey | A 'give it a go' breast-feeding culture and early cessation among low-income mothers[END_REF] Mothers were prepared to give breastfeeding a 'go', but there was a strong expectation of difficulties and even failure. Expertise and confidence with bottle feeding was more widespread amongst support networks of family and friends and women relied strongly on this advice in the transition from hospital to home. A recent cross -cultural study of breastfeeding initiation comparing ethnic groups in the UK and the USA also concluded that the relationship between socioeconomic status and infant feeding practice is strongly influenced by the cultural/social context. Kelly et al [START_REF] Kelly | Racial/ethnic differences in breastfeeding initiation and continuation in the United Kingdom and comparison with findings in the United States[END_REF] found that the advantage shared by black and other ethnic minority groups in the UK with regard to breastfeeding was not observed in the US, where the lowest rate of breastfeeding was found amongst disadvantaged non-Hispanic Black mothers.

Similar conclusions can be drawn from the higher immunisation rates for South Asian infants. [START_REF] Mixer | Ethnicity as a correlate of the uptake of the first dose of mumps, measles and rubella vaccine[END_REF][START_REF] Bhopal | Immunization uptake of Glasgow Asian children: paradoxical benefit of communication barriers?[END_REF] For example the higher uptake of MMR vaccine for mothers of South Asian ethnicity has been attributed to trust in the judgment of health professionals, a tradition of belief within communities in the protective effect of child immunisation and language barriers leading to a lack of awareness of the debate about the safety of MMR. [START_REF] Mixer | Ethnicity as a correlate of the uptake of the first dose of mumps, measles and rubella vaccine[END_REF] The health advantage for infants from South Asian ethnic groups could also relate to the selective migration of healthier women of childbearing age whose positive attitudes towards child protection are drawn from practices in their country of origin. [START_REF] Bakeo | Investigating variations in infant mortality in England and Wales by mother's country of birth, 1983-2001[END_REF] South Asian communities in Manchester increased by half from the 1991-2001 Censuses, with the largest increase in the Pakistani ethnic group.[21]

Policy Implications

The advent of the NHS was marked by improvements in maternal and child health that were the consequence of preventative services that reached the poorest people in the UK population. The problem identified in this study is that, particularly for White infants living in deprived inner city areas in the UK, the health effects of socioeconomic disadvantage are compounded by less than optimum adherence to practices that are protective of child health. Reliance on methods of health education and service delivery that are designed for the general population are unlikely to be successful in this context, but the evidence base of interventions developed specifically for disadvantaged populations is sparse and characterised by small scale studies. [START_REF] Prosser | Interventions to reduce child health inequalities by promoting breast feeding in disadvantaged groups[END_REF] In general the evidence that does exist suggests that methods of service delivery that take into account the social and cultural context of mothers living in deprived areas are more effective in increasing preventative practice. For example peer support in the community has been found to increase breastfeeding for low income women and adolescent mothers and non -professional voluntary support to improve immunization uptake in deprived communities. [START_REF] Barnes | Impact of community volunteers on immunization rates of children younger than 2 years[END_REF][START_REF] Johnson | Community mothers' programme: randomised controlled trial of non-professional intervention in parenting[END_REF][START_REF] Whittaker | Lay workers improving the uptake of immunization[END_REF], [START_REF] Prosser | Interventions to reduce child health inequalities by promoting breast feeding in disadvantaged groups[END_REF][START_REF] Chapman | Differential response to breastfeeding peer counseling within a low income, predominantly Latina population[END_REF][START_REF] Dykes | Infant Feeding Initiative: a report evaluating the breastfeeding practice projects 1999-2002[END_REF][START_REF] Fairbank | A systematic review to evaluate the effectiveness of interventions to promote the initiation of breastfeeding[END_REF][START_REF] Baker | Setting Standards for Preventative Services to Reduce Child Health Inequalities in Greater Manchester[END_REF] This study identified ethnic inequalities in uptake of immunisation and breastfeeding initiation that favoured BME groups and that were independent of relative deprivation; this has implications for the targets set by the Department of Health for monitoring change in child health inequalities. The findings of our study suggest that figures for breastfeeding initiation and immunisation are likely to be inflated in deprived areas by the presence of South Asian and Black or Black British ethnic minority groups and changes over time could reflect the mobility of these populations rather than the impact of policy interventions. A similar picture would emerge for rates of immunisation. Coding for ethnic group membership is not as yet common practice in routinely collected Child Health data and the poor quality of record completion has been noted elsewhere. [START_REF] Mixer | Ethnicity as a correlate of the uptake of the first dose of mumps, measles and rubella vaccine[END_REF] And yet these data are essential for monitoring inequalities in uptake of preventative interventions in infancy and early childhood.

In the UK change in child health inequalities is currently monitored using the headline indicator of 'a reduction of at least 10% in the gap in infant mortality between manual groups and the population as a whole by 2010'. [START_REF]Review of the Infant Mortality Health Inequalities PSA Target[END_REF] The extent to which this constitutes a marker of child health advantage and disadvantage that is useful for locality based policies is questionable, particularly in deprived inner city areas with ethnically diverse populations.

10182 100 1341 100 3013 100 504 100 100 2803 100

		No	%	No	%	No	%	No	% No % No	%		%
	Ethnicity	20203 100 11088 51 1461	7 3280 15 541 3		3	3183 15
	Single Parent
	No	15997 87 8436	83 1093 82 2952 98 501 99		99	2403 86
	Yes	2464	13 1746	17 248	18	61	2	3	1	6	1	400	14
	Total 18461 100 Age
	<17	230	1	176	2	25	2	3	0	0	0	1	0	25	1
	17-18	1045	5	764	7	97	7	52	2	0	0	5	1	127	4
	19-25	6976	34 3936	36 521	36 1160 36 134 25		46	928	30
	26-29	4322	22 2011	18 295	20 913	28 209 39		25	730	23
	30-34	4486	23 2421	22 285	20 758	23 138 26		20	755	24
	35+	2971	15 1728	16 229	16 354	11 51 10 50	8	559	18
	Total	20030 100

11036 100 1452 100 3240 100 532 100 100 3124 100 Parity

	Primiparous	8027	42 4634	43 639	46 1094 35 278 54	30	1191 46
	One previous birth	5865	30 3318	31 420	30 948	30 146 28	29	852	33
	Two previous births	3009	15 1536	14 193	14 621	20 59 11	22	461	18
	Three + previous births	2519	13 1226	11 148	11 482	15 36	7	19	510	20
	Total

19420 100 10714 100 1400 100 3145 100 519 100 100 3014 117 Deprivation*

	Least deprived quartile	5505	27 3170	29 311	21 1367 42 171 32	32	281	9
	2nd quartile	3747	19 1775	16 258	18 961	29 122 23	22	488	15
	3rd quartile	2285	11 1355	12 153	10 338	10 71 13	17	255	8
	Most deprived quartile	8666	43 4788	43 739	51 614	19 177 33	29	2159 68
	Total	20203 100

11088 100 1461 100 3280 100 541 100 100 3183 100

	*2nd and 3rd quartile scores have been merged for the purposes of statistical analysis into one
	interquartile' category

Table 1 : Maternal characteristics of the sample by ethnic group

 1

		Black or
		Black
	Indian Bangladeshi	British

Table 2 : Immunisation uptake and breastfeeding for White, Mixed, South Asian and Black/Black British ethnic groups

 2

	Ethnic Groups		Uptake			Odds Ratios
					Unadjusted Odds		Adjusted ** Odds
		N	Total	(%)	Ratio (95% CI)	P	Ratio (95% CI)	P
	Uptake of MMR vaccine*
	All ethnic groups	10,106	11,261	(90)
	White	5,600	6,392	(88)	1		1
	Mixed	759	846	(90)	1.23 (0.98-1.56)	0.08	1.28 (0.99-1.64)	0.06
	Pakistani	1,716	1,802	(95)	2.82 (2.24-3.55)	<.001	2.84 (2.20-3.65)	<.001
	Indian	281	294	(96)	3.06 (1.74-5.36)	<.001	3.00 (1.58-5.67)	0.002
	Bangladeshi	354	373	(95)	2.64 (1.65-4.21)	<.001	2.71 (1.67-4.40)	<.001
	Black/Black British	1,396	1,554	(90)	1.25 (1.04-1.50)	0.02	1.52 (1.24-1.86)	<.001
	Uptake of Triple Vaccine
	All ethnic groups	18,818	20,203	(93)
	White	10,236	11,088	(92)	1		1
	Mixed	1,330	1,461	(91)	0.85 (0.70-1.03)	0.09	0.80 (0.65-0.99)	0.04
	Pakistani	3,121	3,280	(95)	1.63 (1.37-1.94)	<.001	1.78 (1.47-2.16)	<.001
	Indian	511	541	(94)	1.42 (0.98-2.06)	0.07	1.22 (0.81-1.83)	0.35
	Bangladeshi	626	650	(96)	2.17 (1.44-3.28)	<.001	2.60 (1.68-4.02)	<.001
	Black/Black British	2,994	3,183	(94)	1.32 (1.12-1.55)	<.001	1.58 (1.31-1.91)	<.001
	Breastfeeding at 2 weeks
	postpartum
	All ethnic groups	5,008	8,946	(58)
	White	1,700	4,677	(36)	1		1
	Mixed	363	631	(58)	2.37 (2.00-2.81)	<.001	3.00 (2.49-3.65)	<.001
	Pakistani	1,097	1,528	(72)	4.46 (3.93-5.06)	<.001	4.53 (3.94-5.23)	<.001
	Indian	211	248	(85)	9.99 (7.01-14.2)	<.001	8.40 (5.78-12.2)	<.001
	Bangladeshi	247	294	(84)	9.20 (6.70-12.6)	<.001	10.51 (7.58-14.5)	<.001
	Black/Black British	1,390	1,568	(89)	13.6 (11.6-16.2)	<.001	16.35 (13.5-19.7)	<.001
	* Take up of MMR restricted to the number of children who were at least 2 years old at the time the database was compiled
	** Adjusted for area level of deprivation, parity, parenthood status and age

Table 3 : Independent predictors of immunisation and breastfeeding within White, Pakistani and Black or Black British Ethnic groups.

 3

	White	Pakistani	Black/Black British
	OR (95% CIs)	OR (95% CIs)	OR (95% CIs)

ACKNOWLEDGEMENTS

Contributors: Deborah Baker was responsible for designing the study and for writing the paper. Adam Garrow cleaned and analysed data from the Child Health System, performed the statistical analyses, missing values analyses and wrote up the methods section of the paper. Chris Shiels assisted with the logistic regression analysis. All authors had full access to all the data and read and revised draft versions of the manuscript. DB is guarantor.

We thank all those working with the Child Health System database in Manchester for their kind contribution to this project. Particular thanks are due to Jeanette Beckett and Paul Westhead. We also thank the three reviewers of this article for their helpful and constructive comments.

"The Corresponding Author has the right to grant on behalf of all authors and does grant on behalf of all authors, an exclusive licence (or non exclusive for government employees) on a worldwide basis to the BMJ Publishing Group Ltd and its Licensees to permit this article (if accepted) to be published in BMJ editions and any other

COMPETING INTERESTS: "All authors declare that the answer to the questions on your competing interest form (http://bmj.com/cgi/content/full/317/7154/291/DC1) are all No and therefore have nothing to declare".

WHAT THIS PAPER ADDS

What is already known on the subject

• There is acceptance that inequalities in health by ethnic group are, in general, explained by relative deprivation. • A review of the evidence focusing on interventions that are protective of health in early childhood suggests the independent effect of ethnicity as a source of inequality.

What this study adds

• This study compares three interventions -uptake of MMR, Triple Vaccine and breastfeeding -across ethnic groups taking into account the potential confounding effect of relative deprivation. • It demonstrates that only for the White ethnic group is lower uptake of immunisation and breastfeeding consistently associated with deprivation and that for this group the likelihood of uptake of all three interventions decreases with increasing parity. • Equitable provision of preventative services in early childhood needs to take account of both ethnicity and deprivation as sources of inequality.


Emily Grundy
email: emily.grundy@lshtm.ac.uk

What this study adds

In England and Wales risks of moving to an institution in 1991-2001 were lower, and subsequent excess mortality of those making such a move, higher than in earlier decades. This suggests that policy reforms introduced in the 1990s led to a reduction in the use of institutional care and greater targeting of this resource on the most frail. Chances of living with relatives also show a downward trend suggesting other influences driving increased residential independence of older people. Differences in the mortality of older people living with relatives and those living in institutions indicates that the health status of these populations is not equivalent.

Introduction

Over the past three decades living alone or just with a spouse has become increasingly prevalent and living with relatives increasingly unusual for older people in most of Europe and North America. [START_REF] Grundy | The living arrangements of elderly people[END_REF][START_REF] Pampel | Trends in living alone among the elderly in Europe[END_REF][START_REF] Tomassini | Living arrangements among older people: an overview of trends in Europe and the USA[END_REF] Possible explanations encompass positive developments including higher incomes, better health, and improvements in housing, assistive technology, transport and communications. Such changes may have enabled more older people to meet preferences for residential independence and made it easier to provide extra-household help to those who cannot manage unaided. [START_REF] Mcgarry | Social security, economic growth, and the rise in elderly widows' independence in the twentieth century[END_REF][START_REF] Hoenig | Does assistive technology substitute for personal assistance among the disabled elderly?[END_REF][START_REF] Freedman | Trends in the use of assistive technology and personal care for late-life disability, 1992-2001[END_REF] Other interpretations have emphasised constraints, including possible declines in family support. [START_REF] Goldscheider | New families, no families?[END_REF][START_REF] Wolf | Population change: friend or foe of the chronic care system?[END_REF] The proportion of older people with a spouse or child alive has increased in recent decades in England and Wales and other European countries because of historic changes in family formation patterns and mortality declines leading to later ages at widowhood. [9-11] However, the ability or willingness of younger generations to provide care, including co-resident care, for disabled older relatives (and older people's care preferences), may have fallen as a result of social changes including increases in the proportions of women working full time and greater individualism. [START_REF] Doty | The impact of female caregivers' employment status on patterns of formal and informal eldercare[END_REF][START_REF] Johnson | Family support of the elderly and female labour supply: trade-offs among caregiving, financial transfers and work: evidence from the US Health and Retirement Survey[END_REF][START_REF] Daatland | Intergenerational solidarity and the family-welfare state balance[END_REF] As family care is so important for older people with disabilities, any change in the availability of such support has implications for the demand for formally provided services, including institutional care. [15] Conversely, changes in the supply of formal care have implications for potential family supporters and long-term care policies are an important potential influence on residential choices and constraints among older people with disabilities. [START_REF] Ogawa | Shifting the cost of caring for the elderly back to families in Japan: Will it work?[END_REF]17] Long-term care policy and provision in England and Wales 1Long-term care provision for older people in England and Wales still bears the hallmark of the post World War II legislation which established the modern British welfare state. This included a requirement for local authorities to provide residential care and an empowerment to financially support residents in homes provided in the private (for profit) and voluntary (not-for-profit) sectors. Long-term hospital care became the responsibility of the National Health Service (NHS) and was free of charge, although local authority accommodation was from its inception means tested. Subsequent legislation incrementally extended and changed the relative responsibilities of the NHS and local authorities for providing other long-term care services, such as home help, meals, and day care. [START_REF] Means | From Poor Law to Community Care: The development of services for elderly people[END_REF][START_REF] Wanless | Securing good care for older people: taking a long-term view[END_REF] The past three decades have seen significant changes in both organisation and availability of long-term care services. After a period of expansion in the immediate post war decades, in the later 1970s and 1980s financial constraints meant that the provision of local authority paid-for residential and domiciliary long-term care services failed to keep pace with increases in the size of the older old population. [START_REF] Grundy | Falling rate of provision of residential care for the elderly[END_REF][START_REF] Grundy | Community care for the elderly 1976-1984[END_REF] Long-term hospital provision reduced substantially from the 1980s (in part associated with reductions in lengths of stay) with further large reductions more recently. [START_REF] Brocklehurst | Long-term care-United Kingdom[END_REF][START_REF] Laing | Care of the elderly -UK Market Report[END_REF] Secondly, administrative changes in regulations on payment of supplements to those on low incomes introduced in 1980 made 'board and lodging' payments available to low income older people entering private and voluntary residential or nursing homes resulting in a huge expansion in provision and a ten-fold increase in (Central) Government expenditure on these supplements during the 1980s. [START_REF] Laing | Financing long-term care: the crucial debate[END_REF] This increased provision was associated with changes in the balance between institutional care and co-residence with relatives. Research using data from the Office for National Statistics Longitudinal Study (ONS LS) showed that, after controlling for age, marital status and housing tenure, the risk of transition from a private household to an institution was some 33-52% higher in 1981-91 than in 1971-81 while rates of transition to live with relatives declined. [START_REF] Grundy | Trends in, and transitions to, institutional residence among older people in England and Wales, 1971 to 1991[END_REF] Subsequently, attempts to rein back escalating costs and re-emphasise community care led to the introduction of the NHS and Community Care Act in 1990 (fully implemented in 1993). This returned to local authorities responsibility for arranging and funding (on a means tested basis) long-term care in residential and nursing homes. This legislation introduced a requirement for an assessment of older people moving into residential or nursing home care (apart from self-funders) and the targeting of home care resources on those most at risk of such a move. These policies appear to have had some effect in that admission rates levelled off during the 1990s [START_REF] Brocklehurst | Long-term care-United Kingdom[END_REF][START_REF] Laing | Care of the elderly -UK Market Report[END_REF] and domiciliary services have become focussed on a smaller proportion of older people who receive larger amounts of support. [START_REF] Pickard | The decline of intensive intergenerational care of older people in Great Britain, 1985-1995[END_REF] It is not known whether these changes have influenced rates of co-residence with relatives. There is also a lack of national information on whether the requirement for medical assessment prior to institutional admission led to a change in the health status of older people entering institutional care and their subsequent mortality.

Aims

The aims of this study were to identify possible effects of the NHS and Community Care Act and subsequent reforms firstly by analysing changes in the proportions of older people living with relatives and moving to institutional care in three successive decades in England and Wales, and secondly by comparing the subsequent mortality of older people living with relatives; those living 'independently'; and those living in institutions.

Data and Methods

I employ a cross sequential design to examine the household circumstances of older people at the beginning and end of three successive decades (1971-81, 1981-91 and 1991-2001) and contrast three end-of-decade living arrangements -residence in an institution; co-residence with relatives; and living alone or just with a spouse-, using multinomial logistic regression. Poisson regression was used to model differentials in mortality for each decade-specific sample during the subsequent five years (the precise follow-up period was 4.5 years, this was chosen because of slight variations in the dates of the 1981, 1991 and 2001 Censuses which would have meant variable lengths of followup if end of the calendar year five years post census had been chosen as the end point).

Data are drawn from the ONS LS, a record linkage study of approximately 1% of the population England and Wales. The sample was initially drawn from the 1971 Census and now includes information from subsequent censuses and from vital registration. The LS is a multi-cohort moving sample; members are lost through emigration and death and recruited through immigration and birth. The data set thus remains representative and may be used for cross sectional and cross sequential, as well as longitudinal, analyses. Strengths of the data include large sample size, low non-response and attrition bias (as census coverage is good and rates of linkage high); and inclusion of the institutional population. [START_REF] Blackwell | Longitudinal Study 1971-2001: completeness of census linkage[END_REF] Information from the census records of people in LS members' households is also available, although records for co-residents are not linked over time.

The sample for each decade comprised people aged 65 and over at the start of the decade who were still alive and in the sample ten years later. Those in the analysis sample for later decades thus include survivors from preceding decades together with those attaining the age of 65 in the interim and new recruits to the study. Initial modelling was undertaken separately for each decade but results reported here are for the pooled sample with decade as a co-variate. All analyses were undertaken for men and women separately, because of well established gender differences in living arrangements and mortality, and exclude very small proportions with missing information on relevant variables. People who at the start of each decade were resident in institutions were also excluded (as the intention was to analyse variations in moves to institutions) as were visitors (such as acute hospital inpatients) at the address of enumeration at start or end of decade.

Measures

Private (non-institutional) households in the census, and so in the LS, are defined as individuals or groups of co-residents who share a dwelling and have common housekeeping. People living in establishments where meals and services are communally provided (including those in schemes in which individual cooking facilities are available to fewer than half the residents) are identified as living in institutions. [START_REF] Blackwell | Longitudinal Study 1971-2001: completeness of census linkage[END_REF] For those in private households two classifications of household/family type were derived referring respectively to living arrangements at the start and end of each decade. These have been described in detail elsewhere [START_REF] Grundy | Socio-demographic variations in moves to institutional care 1991-2001: a record linkage study from England and Wales[END_REF] and use the census definition of a family which counts configurations of parents and never-married children (of any age) as part of the same family but not other combinations of relatives (such as widows living with married children). Start of decade living arrangements were grouped into four categories comprising: people living alone, ('Solitary'); people living with a spouse (or partner) only, ('Couple alone'); those in any other type of family, very largely comprising people living with a spouse and children or lone parents living with a never-married child, ('Family'); people living with relatives or friends not classified as being part of the same family or, less usually, in two family households ('Complex household ').

The 'family' and 'complex household' groups were distinguished in the start-of-decade household classification because in the younger elderly population the former often includes people providing support to co-residents (e.g. 65 year old married men with children still at home), whereas the latter configuration is more likely to arise because of parental needs (e.g. 85 year old widows who have moved to live with a married daughter). However, the proportions in the 'family' category by the end of each decade (when sample members were aged 75 and over) were very small, so they were combined with the complex household group in the end of decade classification. Those living alone or just with a spouse at end of decade were also grouped together but are effectively distinguished in the analysis through inclusion of marital status as a co-variate. The third end of decade group comprised those living in an institution.

Other covariates were age in single years and housing tenure (distinguishing home owners from others) at start of the relevant decade and marital status at end of decade. Housing tenure was included as an indicator of socio-economic status and has been shown in previous studies to be associated with household changes, including moves to institutions. [START_REF] Grundy | Socio-demographic variations in moves to institutional care 1991-2001: a record linkage study from England and Wales[END_REF] There may also be a more direct relationship between housing wealth and entry to institutional care as means testing procedures mean that home owners (and their heirs) have particular disincentives to enter state funded residential care. [START_REF] Hancock | The effect of older people's economic resources on care home entry under the United Kingdom's long-term care financing system[END_REF] Marital status is an important indicator of availability of potential care from a spouse and also of likelihood of availability of a child (as in the cohorts considered few nevermarried people have children). End-of-decade indicators for marital status were chosen as being closer temporally to end-of-decade living arrangements and also to avoid problems with collinearity between initial household type and marital status. Finally, a dummy variable indicating decade of observation was derived.

Results

Table 1 shows the distribution of the three samples by variables used in the analysis. The ageing of the England and Wales population is reflected in larger sample sizes for the later decades and slight increases in mean age. The progressively lower proportion of never-married women in each of the three samples is a result of cohort differences in marriage; as would be expected, a much higher proportion of men than women were married. Changes in housing tenure patterns over time, including the consequences of the 1980 'right to buy' legislation which enabled public sector tenants to buy their homes at discounted prices, are evident in the large increase in the proportion of home owners. The distribution of the three samples by household type at the start and end of the relevant decade is shown in Table 2. The proportions living alone at the start of each decade were much higher for women than men and increased in successive decades, in the case of women from 34% in 1971 to 40% in 1991. The proportions living alone or in a couple only household at the end of each decade also increased, for example from 76% for men in 1981 to 83% in 2001. The proportions living with relatives declined and the proportions in institutions first rose and then fell.

Table 3 presents results from fitting multinomial regression models presented as relative risk ratios (and 95% confidence intervals) for three end of decade contrasts: living in an institution rather than alone or with a spouse; living with relatives (family/complex) rather than alone or in a couple; and living in an institution rather than with relatives. Living in an institution, rather than alone or in a couple, was positively associated with older age, living in the family or complex household types at start of decade, being a tenant at start of decade and with being unmarried, particularly never-married, at end of decade. Living in a family/complex household at end of decade rather than alone or in a couple was positively -and very strongly-associated with being in one of these types of living arrangement at the start of the decade, with older age and with being unmarried (although this association was not as great as in the institution versus solitary/couple contrast), and negatively associated with being a tenant. The third contrast, between living in an institution versus with relatives, showed positive associations with older age, being a tenant and being unmarried and strong negative associations with initially living in a complex or family household and, to a lesser extent, with initially living with a spouse. Relative to 1981-91, risk ratios of being in an institution at the end of the period, rather than in the alone/couple category group, were lower in 1991-2001 and in 1971-81 for both women and men. In 1991-2001 the risk of being in an institution rather than living with relatives was also significantly lower than in 1981-91 for women; among men this risk ratio was also lower in 1991-2001 than in 1981-91 but the difference was not statistically significant, possibly because of reduced statistical power in the smaller samples of men. For both men and women risks of being in an institution rather than living with relatives were nevertheless significantly higher for 1991-2001 than for 1971-81. Finally risks of being in a family/complex household rather than solitary/couple show a downward trend over time for women being significantly higher 1971-81 than 1981-91 but lower in 1991-2001. Among men the ratio for the last decade was lower than for 1981-91, but the difference was not statistically significant. In all periods those living alone or in a couple had the highest survival and those in institutions the lowest. However, whereas the survival of men and women living alone/in a couple or with relatives tended to improve over successive periods, the median survival of those in institutions in 2001 was about six months less than median survival of the 1991 institutional population Results from fully adjusted Poisson regression models of mortality for the 4.5 years following each decade of interest are shown in Table 4. Information on co-variates comes from the same time points as shown in earlier tables so, for example, mortality 2001-05 is shown in relation to housing tenure in 1991; marital status in 2001, and household type in both 1991 and 2001. Residence in an institution at end of decade was associated with elevated mortality risks in the subsequent 4.5 years; this excess mortality was greatest in the most recent period. Thus the IRR 2001-05 for residents in an institution in 2001 was 2.85 (95% C.I. 2.68-3.03) for women and 2.80 (2.53-3.09) for men; comparable estimates for 1991-95 were 1.97 (1.86-2.09) for women and 2.19 (1.98-2.29) for men; the extent of excess mortality risk in 1981-85 was lower still. Women living with relatives had higher mortality than the reference category of those living alone or in a couple, but the excess risk was much less marked than for those in institutions and risk ratios were lower, for men and women, than for institutional residents. Household type at start of each relevant decade had no association with mortality 10-15 years later. Analysis of models including interaction terms (not shown) did, however, indicate raised mortality risks for men and women changing from living in a couple in1991 to an institution in 2001; (men: 1.31, 1.06-1.61 p<0.05; women 1.1.6, 1.02-1.32 p<0.05); and for women moving from living in the other family category to living in an a institution 1991-2001 (1.43, 1.09-1.88 p<0.01) and men changing from living in a couple to living in an institution 1981-91 (1.31, 1.04-1.67 p<0.05).

Mortality by end of decade living arrangement

Results also show higher mortality for widowed and divorced compared with married men in 1981-85 and for both widowed and divorced men and women in 2001-05 and a positive association between older age and mortality, which was greater in the 2000-05 period than earlier. There were no significant associations between being never-married and mortality once household type at two time points was controlled for, as shown in the table. In 2001-05 the mortality of men who had been tenants in 1991 was 25% higher than that of 1991 owner occupiers; among women this excess was 20%. Mortality 1991-5 was also raised for men who had been tenants in 1981.

Discussion

For men and women risks of moving to an institution 1991-2001 relative to chances of living alone or in a couple were lower than in 1981-1991. For women risks of moving to an institution rather than living with relatives were also significantly lower 1991-2001 than in 1981-91, although for both men and women they were higher than in 1971-81. Chances of living with relatives rather than alone or with a spouse were lower in 1991-2001 than 1971-81 or, for women, in 1981-91. Excess mortality of the institutional population was greatest in 2001-05. Among women, those living with relatives also had raised mortality risks, but these were considerably less than the risks for those in institutions suggesting that these populations are not equivalent in terms of health status. Housing tenure was more strongly associated with mortality risk in 2001-05 than in earlier periods probably because tenants by then comprised a smaller and relatively more disadvantaged group. It is not clear why age was more strongly associated with mortality in 2001-05 than in previous periods but investigation using data from the whole England and Wales population showed a similar increase in the strength of the association between age and mortality, a change which merits further analysis.

These results are based on a large nationally representative data set that is minimally affected by non response and attrition and allows comparisons over three decades. However the long intervals between census data capture points makes it impossible to pinpoint dates of household transitions; we do not know, for example, whether changes in household type between 1991 and 2001 pre or post dated implementation of the NHS Community Care Act, although given the high mortality of residents in institutions it is unlikely that many of those in institutional care in 2001 had been there for more than seven years. The lack of detailed information on health, family support from outside the household or use of services presents a further limitation, particularly in this design as the analysis was restricted to variables included in all the censuses of interest. Studies including more detailed information have been able to identify factors predictive of entry to institutions more precisely and in particular have identified the importance of conditions such as dementia and the preventive role of extra-household support from relatives and formal services. [30-33] However, in many cases these more detailed findings have been restricted to local populations and do not allow consideration of changes over time [30,[START_REF] Crawford | Comparison of residential and nursing home care before and after the 1991 community care policy[END_REF] and as yet we lack data sets which combine detail with size and availability of data for different time periods.

The findings suggest that policy changes influence the residential choices of older people and that more specifically, the NHS and Community Care Act and other reforms in the 1990s led to a reduction in institutional admission and that those who were admitted had a poorer health status overall than people entering institutional care in previous decades. Among women the balance between institutional and co-resident family care also shifted slightly towards the latter in the 1991-2001 decade (although not back to the 1971-81 level), but also towards living alone (or with a spouse). The targeting of community services on those deemed most at risk of institutional admission, which included increases in hours of help provided for this group, [START_REF] Pickard | The decline of intensive intergenerational care of older people in Great Britain, 1985-1995[END_REF] may have enabled more people to remain in the community for longer, even if living alone or just with a spouse. Overall chances of living with relatives were lower in 1991-2001 than in previous decades suggesting other influences driving a continued trend towards increased residential independence among older people. International comparisons would improve our ability to identify these differing drivers, and perhaps identify more clearly the effects of policy. Finally, despite the high mortality of residents in institutions, it is noteworthy that 36% of women and 26% of men in institutional care in 2001 were still alive three years later, a finding relevant to both families and service providers planning financing of care.

Figure

  Figure 1a and b show the survival of men and women for the four and a half years following each of the 1981, 1991 and 2001 Censuses by living arrangement at that point.In all periods those living alone or in a couple had the highest survival and those in institutions the lowest. However, whereas the survival of men and women living alone/in a couple or with relatives tended to improve over successive periods, the median survival of those in institutions in 2001 was about six months less than median survival of the 1991 institutional population Results from fully adjusted Poisson regression models of mortality for the 4.5 years following each decade of interest are shown in Table4. Information on co-variates comes from the same time points as shown in earlier tables so, for example, mortality 2001-05 is shown in relation to housing tenure in 1991; marital status in 2001, and household type in both 1991 and 2001. Residence in an institution at end of decade was associated with elevated mortality risks in the subsequent 4.5 years; this excess mortality was greatest in the most recent period. Thus the IRR 2001-05 for residents in an institution in 2001 was 2.85 (95% C.I. 2.68-3.03) for women and 2.80 (2.53-3.09) for men; comparable estimates for 1991-95 were 1.97 (1.86-2.09) for women and 2.19 (1.98-2.29) for men; the extent of excess mortality risk in 1981-85 was lower still. Women living with relatives had higher mortality than the reference category of those living alone or in a couple, but the excess risk was much less marked than for those in institutions and risk ratios were lower, for men and women, than

Figure 1 .

 1 Figure 1. Survival (proportion surviving) by living arrangement at census, 1981-5, 1991-5 and 2001-05.

Table 1 . Distribution of 1971-81, 1981-91 and 1991-2001 samples by socio- demographic characteristics.

 1

		1971-81		1981-91		1991-2001
		Men	Women Men	Women Men	Women
	Mean age 1 (SD)	69.5	70.6	70.0	71.2	70.0	71.2
		(4.1)	(4.7)	(4.3)	(5.0)	(4.4)	(5.0)
	Owner occupier 1 (%) 58.4	52.7	61.3	56.0	72.6	65.8
	Tenant 1 (%)	41.6	47.7	38.7	44.0	27.4	34.2
	Married 2 (%)	61.2	20.2	63.8	23.8	63.6	25.9
	Never-married 2 (%)	6.1	13.0	5.4	10.1	5.5	7.0
	Wid./div/ 2 (%)	32.7	66.8	30.8	66.2	30.9	67.1
	N	8,482 17,259	10,971 20,978	13,587	23,107
	1 At start of decade; 2 at end of decade.
	Source: Analysis of ONS Longitudinal Study data.

Table 3 . Results from multinomial regression models of transitions between household types 1971-81; 1981-91 and 1991-2001, men and women aged 65 and over at start of decade. Household type at end of decade Institution vs. solitary/couple alone Family/complex vs. solitary/couple alone

 3

	Institution vs.
	Family/complex

Table 4 .

 4 Results from Poisson regression analysis of mortality by household type at start and end of relevant decade, 1981-85; 1991-95 and 2001-05, men and women aged 65 and over at start of decade.

		1981-85		1991-95	2001-05
	MEN	IRR	95% C.I.	IRR	95% C.I. IRR	95% C.I.
	Age	1.04***	1.04-1.05	1.03***	1.03-1.04 1.08***	1.08-1.09
	Owner occupier (Ref.) 1.00		1.00	1.00
	Tenant	1.04	0.99-1.10	1.08**	1.03-1.14 1.25***	1.18-1.33
	Household type at start of decade
	Solitary (Ref.)	1.00		1.00	1.00
	Couple alone	1.04	0.94-1.15	1.02	0.93-1.11 0.98	0.90-1.08
	Family	1.04	0.92-1.17	0.97	0.87-1.09 1.00	0.88-1.13
	Complex	1.02	0.90-1.16	0.98	0.86-1.10 0.94	0.81-1.09
	Household type at end of decade
	Solitary/couple (Ref.)	1.00		1.00	1.00
	Family/complex	1.08	0.99-1.17	1.07	0.98-1.16 1.10	1.00-1.21
	Institution	1.91***	1.68-2.18	2.19***	1.98-2.29 2.80***	2.53-3.09
	Marital status
	Married (Ref.)	1.00		1.00	1.00
	Single	1.04	0.91-1.19	1.00	0.88-1.15 1.11	0.96-1.28
	Widowed/divorced	1.10**	1.03-1.18	1.02	0.96-1.08 1.11*	1.03-1.18
	Number of deaths	1162		1230	1204
	WOMEN
	Age	1.04***	1.04-1.04	1.04***	1.04-1.04 1.09***	1.09-1.09
	Owner occupier (Ref.) 1.00		1.00	1.00
	Tenant	1.04	1.00-1.08	1.04	1.00-1.08 1.20***	1.15-1.26
	Household type at start of decade
	Solitary (Ref.)	1.00		1.00	1.00
	Couple alone	0.99	0.93-1.04	0.97	0.93-1.02 1.03	0.97-1.10
	Family	0.93	0.87-1.01	0.96	0.89-1.04 0.99	0.90-1.08
	Complex	0.98	0.91-1.05	1.01	0.95-1.09 1.03	0.94-1.13
	Household type at end of decade
	Solitary/couple (Ref.)	1.00		1.00	1.00
	Family/complex	1.15***	1.08-1.22	1.10**	1.04-1.17 1.25***	1.15-1.35
	Institution	1.85***	1.72-1.99	1.97***	1.86-2.09 2.85***	2.68-3.03
	Marital status
	Married (Ref.)	1.00		1.00	1.00
	Single	0.98	0.90-1.07	0.97	0.89-1.05 1.04	0.93-1.16
	Widowed/divorced	0.99	0.93-1.05	1.02	0.97-1.08 1.08*	1.00-1.15
	Number of deaths	1596		1705	1641
	Source: Analysis of ONS Longitudinal Study data.
	* p<0.05; ** p<0.01; ***p<0.001

There are a number of differences in long-term care policies in the constituent countries of the UK (England & Wales, Scotland, Northern Ireland) which are not discussed here.

Research reported here was supported by the UK Economic and Social Research Council (Award Ref. RES-339-25-0002). The permission of the Office for National Statistics to use the Longitudinal Study is gratefully acknowledged, as is the help provided by Christopher Marshall and other staff of the Centre for Longitudinal Study Information & User Support (CeLSIUS). CeLSIUS is supported by the ESRC Census of Population Programme (Award Ref: RES-348-25-0004). Census output is Crown copyright and is reproduced with the permission of the Controller of HMSO.

Conflict of interests:None

The corresponding author has the right to grant on behalf of all authors and does grant on behalf of all authors an exclusive license on a worldwide basis to the BMJ Publishing Group Ltd and its licensees to permit this article (if accepted) to be published in JECH editions and any other BMJPGL products to exploit all subsidiary rights, as set out in our license (http:/jech.bmj.com.ifora.licence.pdf).

What this paper adds What is already known

There have been major changes both in the living arrangements of older people and in long-term care policy and provision in England and Wales, and many other European countries. Changes in the availability of co-resident family care for older people has implications for demand for formally provided care, conversely long term care policies are an important potential influence on the residential choices of disabled older people.


Armstrong Bg

Chalabi Z Fenn

Hajat S Kovats

Ben Armstrong
email: ben.armstrong@lshtm.ac.uk

The association of mortality with high temperatures in a temperate climate: England and Wales

Keywords: total mortality, time series, climate, heat

What is already known:

It is well known that high ambient temperatures are associated with increased mortality, even in temperate climates, but some important details remain unclear. In particular how heat-mortality associations (for example slopes and thresholds) vary by climate has previously been considered only qualitatively..

What this paper adds

English regions with hotter climates had higher threshold temperatures at which adverse effects started. Specifically, effects were well approximated by threshold-linear models that can be predicted from the region's climate (93 rd centile and mean summer temperature). On average mortality increased by 2.1% per degree Celsius above the 93 rd centile, but the increment was higher in hotter regions. This suggests partial adaption to hotter climates, relevant to estimating impacts from future climate change.

INTRODUCTION

It is firmly established across many countries and areas that mortality increases at times of ambient temperatures that are unusually high for the area [START_REF] Baccini | Heat Effects on Mortality in 15 European Cities[END_REF][START_REF] Basu | Relation between elevated ambient temperature and mortality: a review of the epidemiologic evidence[END_REF]. It is also widely found that the nature of the temperature-mortality relationship varies between places [START_REF] Baccini | Heat Effects on Mortality in 15 European Cities[END_REF][START_REF] Anderson | Weather-related mortality: how heat, cold, and heat waves affect mortality in the United States[END_REF][START_REF] Curriero | Temperature and mortality in eleven cities of the eastern United States[END_REF][START_REF] Mcmichael | International study of temperature, heat and urban mortality: the 'ISOTHURM' project[END_REF]. In particular adverse effects of heat begin at lower temperatures in climates with lower summer temperatures, suggesting that populations adapt to their climate. [START_REF]Heatwave Plan[END_REF] However, details of temperature-mortality associations (eg slope and threshold) need to be freshly identified for each place -a challenge for estimating burden of deaths due to heat and for setting alarm thresholds in heat watch warning systems. We have sought in this paper to fill this gap by proposing model in which the temperature-mortality association can be deduced from a region's climate. This has not previously been attempted. Specifically, we characterized population temperature-mortality associations in summer in the ten government regions of England and Wales in the period 1993-2006. In particular, we aimed to identify climatic determinants which would explain any variation in these associations between regions and so allow parsimonious models that covered all of England and Wales. Downstream uses for this research include informing the Heatwave Plan for England [START_REF]Heatwave Plan[END_REF], a warning system set up in 2004 by the British Government, and the estimation of impacts of climate change. We also sought to extend methodology for characterising the associations between mortality and temperature across populations with different climates, to facilitate similar research in other parts of the world.

METHODS.

We carried out an ecological time-series regression analysis of daily counts of all-cause mortality and measures of ambient temperature between 1993 and 2006 in the ten government regions of England and Wales [START_REF]Map of Government Official Regions Office for National Statistics[END_REF].

Individual death record data were obtained from the ONS, including date of death and postcode of residence at time of death. The postcode was used to divide deaths by government region and date to make daily series of counts for each region. Here we consider counts of all deaths (including those from external causes).

We downloaded from the British Atmospheric Data Centre [8] all available monitoring station data on minimum and maximum dry-bulb temperature, from which the daily mean was estimated. From 9 am and 3pm dewpoint measurements we also estimated daily mean relative humidity and daily mean and maximum apparent temperature, an index designed to reflect the combined effect of humidity and temperature on heat stress [START_REF] Basu | A multicounty analysis identifying the populations vulnerable to mortality associated with high ambient temperature in California[END_REF]. The data were processed, imputing missing values the method of the AIRGENE study [START_REF] Ruckerl | Air pollution and inflammation (interleukin-6, Creactive protein, fibrinogen) in myocardial infarction survivors[END_REF], to obtain population-weighted mean daily series for each region Further details are given in on-line annex 1.

The temperature series were highly correlated within regions (mean r=0.95, range 0.94-0.98) and station means varied little within region (mean SD = 0.7 o C range 0.3-1.1). Correlation between RH within regions was lower, on average 0.84.

We did not attempt to estimate daily pollution series because variation of daily series from monitors within regions (typically comprising several different urban and rural areas) was expected to be too high to provide meaningful single regional series.

Statistical model

Given the focus in these analyses on heat effects, to avoid complexities of simultaneous modelling cold effects we restricted analyses to summer, defined here as June-September (the hottest four months). Separate models were fit to each region. The region-specific models followed a fairly standard form for season-specific time series regression [START_REF] Baccini | Heat Effects on Mortality in 15 European Cities[END_REF][START_REF] Hajat | Impact of high temperatures on mortality: is there an added heat wave effect?[END_REF]. Our strategy was to fit and report a core confounder model decided a priori and explore the adequacy of this model and sensitivity of conclusions from fitting it through supplementary analyses. The models are summarised here and described formally in on-line annex 2.

All models assumed Poisson variation with scale overdispersion and residual auto-regression. The core confounder model comprised:

1. Within-summer seasonal variation: 4 degrees of freedom (df) natural cubic splines (NCS) in day-of-year (same for all years) 2. Trend in time over 11 years: linear and quadratic terms in time 3. 6 day-of week indicators 4. Relative humidity:3 df NCS for mean time lags 0-1 day (lag 0-1). For sensitivity analyses (because routine incorporation was cumbersome), autoregression was allowed for using one of the lagged residual approach described by Brumback [START_REF] Brumback | Transitional regression models, with application to environmental time series[END_REF]. Terms were included to order three, which was the highest order with absolute partial autocorrelation of residuals above 0.05 in any region.

To describe region-specific temperature-mortality associations we initially used natural cubic splines fit to the daily maximum temperature over lags 0-1, essentially equivalent to a distributed lag model constrained to have equal parameters for lags 0 and 1 [START_REF] Armstrong | Models for the relationship between ambient temperature and daily mortality[END_REF]. Maximum temperature was chosen as the index providing best fit in exploratory analyses. Focus on lag 0-1 was chosen based on previous work in the UK [START_REF] Hajat | Mortality Displacement of Heat-Related Deaths: A Comparison of Delhi[END_REF][START_REF] Pattenden | Mortality and temperature in Sofia and London[END_REF]. We used default (equal quantile) knot placing and 4df -the number that maximised overall Akaike's Information Criterion (AIC). Splines were graphed as predicted mortality relative to that at the temperature at which mortality was minimum [START_REF] Cao | An S-plus function to calculate relative risks for regression models using natural splines[END_REF]. From the fitted spline models we also derived estimated temperature-specific slopes from the analytic first derivative of the spline basis terms.

Following description using the spline curves we fitted the simpler linear threshold (hockey-stick) model [START_REF] Armstrong | Models for the relationship between ambient temperature and daily mortality[END_REF], by which mortality is not related to temperatures below a "threshold" and log-linearly related to temperatures above the threshold. Separate thresholds were initially fit to all regions. Distribution of thresholds and slopes across regions was then explored in a second stage analysis, graphically and using standard meta-analytic methods allowing for random effects between regions [START_REF] Sutton | Methods for meta-analysis in medical research[END_REF]. As will be described in the results section patterns of thresholds and slopes across regions suggested more parsimonious models with thresholds and slopes determined by regional climate.

In the threshold model deaths attributable to heat were estimated using the method of Bruzzi [START_REF] Bruzzi | Estimating the population attributable risk for multiple risk factors using case-control data[END_REF]. First daily attributable deaths were calculated from the number of deaths and the model-predicted relative risk due to heat on each day, then these were summed over all days, and divided by the total number of deaths.

To aid comparison of models, and in particular assess adequacy of the linear-threshold model we computed deviance and AIC corrected for over-dispersion (deviance+2pφ, where p=number of parameters and φ is the estimated scale overdispersion) as well as overdispersion and various summaries of partial autocorrelation of residuals. Sensitivity analyses were carried out to explore robustness of key model features to model assumptions and the use of alternative temperature indices (minimum, mean, apparent).

RESULTS

There were on average 145 deaths per day in a region ranging from 72 in the NE to 200 in the SE. Figure 1 shows the regional temperature distributions (also summarized in Table 1) and fitted spline curves for temperature-mortality associations. In these and subsequent graphs and tables, the regions are shown in order of their mean summer daily maximum temperatures (which ranged from 18.4 o C (NE) to 21.8 o C (London)).

Table 1 The conurbations: summer temperature distributions, number of deaths, increments in relative risk per degree under the linear-threshold model, and attributable deaths. Notes: 1. 93rd centile of year-round two-day mean temperature distribution. 2. DerSimonian and Laird estimate [START_REF] Dersimonian | Meta-analysis in clinical trials[END_REF] Clear evidence of increased mortality at the highest temperatures was apparent for all regions except the cool North-East. The final panel in Figure 1 shows splines for all regions super-imposed on the same graph. Although not clearly discernable from the figure, the splines show subtle differences between regions beyond that attributable to the regions' different range of temperatures,. more clearly apparent if linear-threshold models are fit. For this thresholds were estimated by maximum likelihood (repeated calculation over a grid of values) in each region. The left panel of Figure 2 shows the thresholds and their likelihood-profile confidence interval (CI) by region sorted according to mean summer temperature. Although thresholds are estimated imprecisely, there is a clear pattern of higher thresholds in regions with higher mean temperatures (confirmed significant by meta-regression). The right panel of Figure 2 shows the same graph, but with thresholds expressed as the region-specific percentile of the (year-round) temperature distribution rather than as o C. Heterogeneity remains but is reduced compared to that for thresholds as o C (I 2 =75% Vs 90%) and there is no longer any clear pattern by mean summer temperature. The best-fitting threshold by maximum likelihood (ML), if fixed to be at the same centile in each region, was at the 93 rd centile of (two-day mean) temperatures (95% CI 92 nd -95 th ) .

We used this model (threshold at the regional 93 rd centile) to give a simple quantitative summary of the associations in each region that avoids having to rely on estimates made separately for each region. The region-specific thresholds (93 rd centiles), the estimated slopes (increment in mortality per degree above the threshold) and heat-attributable deaths according to this model are given in Table 1. There was a strong association of slopes of these models with mean summer temperature (Figure 3), strongly statistically significant even if the obvious influential London region was excluded. This suggested possible a parsimonious model in which slopes were a function of the temperature distribution. In particular, we found that slopes were well predicted by regression on the mean summer temperature. (slope= -.1284 +0.0074=-× (mean summer temperature)) . We refer to this model as a climate-based threshold model.

To recapitulate, thresholds are related to the temperature distribution, approximating reasonably to the 93rd percentile in each region. With thresholds so placed, the supra-threshold slopes of the mortality-temperature relationships are still related to temperature, the steepest slopes occurring in the hottest cities. This suggests a climate-based threshold model, by which dependence of mortality on heat can be predicted from the climate of a city.

We additionally explored curvature in the supra-threshold temperature-mortality by adding a quadratic term (details not shown). We found that indeed this addition improved fit significantly.

The direction of curvature varied between regions, but was predominantly positive (higher slopes at higher temperatures).

Table 2 shows deviance and AIC for the various models we have described above and some others, to allow identification of those that fit the data best. To aid comparison deviance and AIC are shown relative to a baseline of the model with all covariates but no temperature terms. Models best fitting for prediction are those with lowest (ie biggest negative) AIC.

- -----------------------------------------------------------------------------------------------------------------------The best fitting spline model (with 4 df) as graphed reduced deviance by 1766.9 and AIC by 1712.2. Some of the simpler models came close to this if number of parameters was penalised (ie by AIC) though did not quite match it. Although among linear-threshold models the one with region-specific thresholds fitted best, that with thresholds fixed at the 93rd centile came quite a close second by AIC. Constraining slopes to their value predicted by summer mean temperature improved fit slightly by AIC, because the loss in deviance terms was outweighed by the reduction in parameters, which are penalised by AIC.

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Comparison based on our prioritised model form (4 df spline for lag 0-1) confirmed daily maximum temperature to fit better than (in order of fit) maximum apparent, mean, mean apparent, or minimum temperatures. Adding spline terms for minimum temperature to the spline for maximum temperature (final block on table 2), improved fit. Replacing the default equal-quantile knots by equal-interval knots also improved fit slightly. Replacing the lag 0-1 mean of maximum temperatures by lag 0 or the lag 0-2 mean however reduced goodness of fit by AIC.

We explored sensitivity of key results to model assumptions. To simplify computing, we confined these to the linear-threshold model with threshold at the 93rd centile, and compared across models the mean slope and degree of heterogeneity between regions as indicated by the range of regionspecific slopes, on the grounds that most substantial sensitivity of other model features would be reflected in sensitivity in this simpler model. Though no model modification substantially changed results, the modification that was associated with greatest changes was that replacing the time terms (linear-quadratic trend and within-year spline terms) with elapsed month indicators (equivalent to a case-cross-over analysis with fixed strata). [START_REF] Farrington | Semiparameteric analysis of case series data[END_REF] In this model the mean supra-threshold slope was greater (2.4% per o C Vs 2.1)) and there was less variation between areas (range 2.6% Vs 3.0%).

The pattern of higher slopes in hotter regions remained significant, however. Using the alternative temperature lags and indices shown on Table 2 changed the unit increments more -unsurprisingly since this changed the scale, but again not the pattern. More detailed results for the sensitivity analyses and model fit statistics are shown in on-line Tables E1 andE2.

DISCUSSION

Our finding that mortality is increased at higher temperatures, even in temperate climates such as England's, is not new. However, we have sought new precision in details of this relationship and patterns in its variation over regions. The following features were prominent in our findings:

• Threshold temperatures for heat effects to commence were at about the same centile temperatures (the 93rd) in all regions. Thus regions with higher summer temperatures on average had higher thresholds. • However regions with higher summer temperatures showed greater increments in mortality per degree above the 93rd centile threshold (slopes), well-characterized by a linear model with mean summer temperature. • Linear-threshold models capture most but not all the association. There was evidence for nonlinear impacts of temperature even above thresholds, with higher slopes at greater temperatures.

These findings suggested a parsimonious model for the pattern of variation in shape of associations between summer temperature and mortality, a climate-based threshold model with threshold following the 93rd centile and supra-threshold slope determined by a linear function of mean summer temperature. Such a parsimonious model has some advantages over region-specific estimates of associations: (i) it "borrows strength" from regions with information to more precisely estimate the association to allow better estimates for regions with less information; (ii) it allows estimation of heat impacts in a region not included in the study, since the relationship is determined by the temperature distribution,. (iii) it allows estimation of heat impacts under climate change, if adaptation is assumed such as that exhibited by the variation across regions (discussed below).

To achieve sufficient simplicity to characterize patterns of heat-mortality association across regions we have however put aside some detail. Perhaps most importantly, we have ignored evidence that the supra-threshold dependence of mortality on temperature is not linear (curved upwards). This can be accommodated in the above model at some complexity cost, for example by adding a quadratic term in supra-threshold temperature as described above.

For the same reason we have also not at this stage incorporated variables which might modify the heat impact. These include time in the season (some studies have shown great increment per degree earlier in the summer [START_REF] Baccini | Heat Effects on Mortality in 15 European Cities[END_REF]21], duration of sustained heat (some studies have shown increased impact for sustained heat), personal vulnerability factors (many studies have shown older people at higher risk, some have shown lower socio-economic status to be at higher risk [START_REF] Hajat | Heat-related and cold-related deaths in England and Wales: who is at risk?[END_REF][START_REF] Medina-Ramon | Extreme temperatures and mortality: assessing effect modification by personal characteristics and specific cause of death in a multi-city case-only analysis[END_REF][START_REF] O'neill | Disparities by race in heat-related mortality in four US cities: the role of air conditioning prevalence[END_REF][START_REF] Schwartz | Who is sensitive to extremes of temperature?: A case-only analysis[END_REF]). Similarly, we have not attempted to distinguish relationships for different causes of death, which are known to exhibit different degrees of sensitivity to heat [START_REF] Baccini | Heat Effects on Mortality in 15 European Cities[END_REF]21,[START_REF] Braga | The effect of weather on respiratory and cardiovascular deaths in 12 U.S. cities[END_REF]. Nor have we further evaluated the extent to which deaths caused by heat are displaced ("harvesting") [START_REF] Hajat | Mortality Displacement of Heat-Related Deaths: A Comparison of Delhi[END_REF][START_REF] Braga | The time course of weather-related deaths[END_REF]. Most of these factors are similar over regions, so lack of allowance for them should not have substantially compromised our goal of characterising differences across regions, Deprivation is known to be greater in the (colder) North, but in England deprivation has not been found to predict vulnerability to heat. [START_REF] Hajat | Heat-related and cold-related deaths in England and Wales: who is at risk?[END_REF] Nevertheless, some residual confounding is likely and could operate differently across regions.

The model we propose is consistent with the observation in several studies that heat thresholds (or minimum mortality temperatures in all-year studies) are higher in hotter places [START_REF] Baccini | Heat Effects on Mortality in 15 European Cities[END_REF][START_REF] Curriero | Temperature and mortality in eleven cities of the eastern United States[END_REF][START_REF] Mcmichael | International study of temperature, heat and urban mortality: the 'ISOTHURM' project[END_REF][START_REF] Braga | The effect of weather on respiratory and cardiovascular deaths in 12 U.S. cities[END_REF][START_REF] Honda | Determination of Optimum Daily Maximum Temperature Using Climate Data[END_REF][START_REF] Kim | High temperature, heat index, and mortality in 6 major cities in South Korea[END_REF]. Honda [START_REF] Honda | Determination of Optimum Daily Maximum Temperature Using Climate Data[END_REF] further found that minimum mortality temperatures across 47 prefectures in Japan were close to 80-85 th centile temperatures. Though the centile is different to the 93rd centile we found, the finding is similar in the sense that the a key feature of the relationship is at a constant centile across regions; the difference in centile could be due to use of all-year data and a spline model rather than summer data and a linear-threshold model. Baccini [START_REF] Baccini | Heat Effects on Mortality in 15 European Cities[END_REF] also found higher heat slopes as well as higher thresholds in the hotter climates in Europe, though the threshold-climate relationship was not explored quantitatively. Anderson also found in US cities that relative risks across centiles of temperature distributions were less heterogeneous than were those for absolute temperatures, though here there was a tendency for lower heat slopes (RR across 99 th to 90 centile) in hotter climates. [START_REF] Anderson | Weather-related mortality: how heat, cold, and heat waves affect mortality in the United States[END_REF] This difference may be due to the substantial use of air conditioning in the US, in particular in its hotter cities.

Of the heat indices considered, maximum temperature fitted the data best, quite closely followed by maximum apparent temperature.. There was some variation between regions in which index fitted best, with London for example favouring daily mean dry-bulb temperature. The relatively poor performance of minimum (night) temperature was somewhat surprising, but should not be taken to imply that these temperatures have no importance. When daily minimum temperature was added to a model addition to with daily maximum temperature, this improved model fit significantly, suggesting that minimum temperature has some importance independently of the maximum. The combination of maximum and minimum temperatures was also found important by Fouillet. [START_REF] Fouillet | A predictive model relating daily fluctuations in summer temperatures and mortality rates[END_REF] We have not in this paper sought to characterise this joint effect. Doing so is not trivial because of the high correlation of the two. There was little evidence that humidity was associated with mortality.

The implication in our climate-based threshold model of centile thresholds is that cities adapt to climate, since the proportion of days affected does not depend on climate. However, the implication of higher incremental risk per degree in hotter cities suggests that the adaptation is partial -there were higher attributable percentages of deaths in hotter cities (table 1). We hypothesized the possibility of using this partial adaptation model to estimate impacts of future climate change. This remains tentative, however. We have observed these patterns cross-sectionally not over time, and over a narrow climate range. Evidence from the USA suggests that factors little operating in England, perhaps air conditioning, can modify the dependence of heat vulnerability across climates.

In summary, we have shown that effects of high daily summer temperatures on mortality in English regions are quite well approximated by threshold-linear models that can be predicted from the region's climate (93rd centile and mean summer temperature). It remains to be seen whether the relationship fits other countries and climates.

Figure 1 Figure 2 Figure 3 .

 123 Figure 1 Temperature-mortality associations in each region (natural cubic splines with 4df; equal quantile knots)

Table 2

 2 Summary of fit of models (lag 01 mean temp) -all regions combined

	Model description*	df	Deviance	AIC
	No temperature	160.0	0.0	0.0
	Spline curves
	3 df	190.0	-1730.3	-1697.4
	4 df	200.0	-1766.9	-1712.2
	5 df	210.0	-1784.3	-1707.4
	6 df	220.0	-1805.8	-1706.7
	Threshold models
	Linear: separate thresholds & slopes	180.0	-1674.8	-1663.7
	Linear: thresh at P93%; separate slopes	171.0	-1629.9	-1638.0
	+quadratic terms	181.0	-1707.6	-1694.7
	Linear: thresh at P93%; slopes constrained	163.0	-1610.1	-1635.6
	Alternative temperature measures (as 4df splines)
	Max (ie core model)	200.0	-1766.9	-1712.2
	Min	200.0	-676.7	-595.2
	Mean	200.0	-1522.4	-1461.9
	Mean apparent temperature	200.0	-1418.7	-1355.7
	Max apparent temperature	200.0	-1742.7	-1687.4
	Further models with 4 df temp spline:
	add 4 df spline for minimum temp	240.0	-1955.0	-1814.7
	Knots placed at equal o C	200.0	-1793.9	-1739.9
	Lag 0	200.0	-1703.6	-1647.0
	Lag 1	200.0	1322.0	1256.8
	Lag 0-2	200.0	-1554.1	-1494.5
	All models have the core covariate terms (day-of-year spline, linear-quadratic long term trend, day-
	of-week indicators, relative humidity spline) unless otherwise specified.
	Deviance and AIC expressed relative to model with not temperature terms (20695.1 and 21089.0
	resp)
	-

Acknowledgements

This study was funded by the UK Medical Research Council.